ADL Profile

Evidence Reviewed as of before: 14-12-2012
Author(s)*: Valérie Poulin; Vanessa Barfod, BA
Editor(s): Annabel McDermott, OT.; Nicol Korner-Bitensky, PhD OT
Expert Reviewer: Carolina Bottari, erg. PhD

Purpose

The ADL Profile is a criterion-referenced measure of independence in everyday activities such as self-care, household management and community activities for individuals with a traumatic brain injury (TBI). The ADL Profile was created by Elisabeth Dutil, Carolina Bottari, Marie Vanier and Céline Gaudreault.

In-Depth Review

Purpose of the measure

The ADL Profile is a criterion-referenced measure of independence in everyday activities in consideration of the important contribution of executive functions for individuals with a traumatic brain injury (TBI) (Canadian Association of Occupational Therapists, 2012; C. Bottari, personal communication, November 6, 2012). The ADL Profile consists of both a performance-based assessment (evaluator’s direct observation of performance) and a questionnaire administered in the form of semi-structured interviews with the person and a significant other (perceptions of person and significant other of person’s functioning). The ADL Profile assesses an individual’s ability to formulate and plan goals for personal and instrumental activities of daily living (PADL and IADL) in interaction with the environment in which they live. Using a task-analysis framework, the individual’s independent performance of ADL tasks is quantitatively analyzed according to 4 executive function operations:

  1. Formulating a goal
  2. Planning
  3. Carrying out the task
  4. Verifying attainment of the initial goal (Bottari et al., 2010b).

The ADL Profile was originally developed for use with patients with traumatic brain injury as an assessment of independence in everyday activities within three environments:

  1. Personal (self-care dimension);
  2. Home (home dimension); and
  3. Community (community dimension).

An exhaustive list of variables was derived from existing ADL tools and were organized as activities, tasks or operations under the three dimensions of personal care, home and community environments, according to Crochart’s (1987) ergonomic model. An expert group of therapists and researchers were consulted to review the refined list of variables and to ensure that all domains related to the concept of ADLs were included in the instrument. A review of the literature on components of ADL assessments also provided support for the experts’ verdict (Dutil et al., 2005).

Available versions

There are no alternate versions of the ADL Profile.

The IADL Profile is a revised version of the ADL Profile (Bottari et al., 2010b) and as such is not included in this review.

Features of the measure

Items:

The ADL Profile consists of 20 PADL and IADL tasks in two parts:

  1. A non-structured, performance-based assessment that comprises observation of 17 tasks; and
  2. A semi-structured interview administered to the patient and his/her significant other that documents 3 tasks (indicated below).

The 20 items relate to (a) personal care; (b) household management; and (c) community activities: Personal care items (6 tasks):

  1. Bathing/showering
  2. Grooming
  3. Toileting
  4. Putting on clothes and shoes
  5. Having a meal
  6. Following his or her diet/taking his or her medication*

Household management (5 tasks):

  1. Preparing a light meal
  2. Preparing a hot meal
  3. Doing daily housecleaning
  4. Doing weekly housecleaning
  5. Doing laundry

Community activities (9 tasks):

  1. Walking or moving outdoors
  2. Using public transportation
  3. Driving a car*
  4. Running errands
  5. Telephoning for information
  6. Paying bill
  7. Using an automatic banking machine
  8. Making a budget
  9. Keeping appointments*

* Three tasks are evaluated by semi-structured interview. Driving is not evaluated per se but certain information is gathered regarding this activity.

Description of tasks:

The client is asked what he/she would normally do at that time of the day and is then given the opportunity to perform the ADL task without assistance from the clinician. The non-structured evaluation enables the clinician to observe deficits relating to executive processes. Accordingly, the clinician informs the client at the onset of the test that he/she will provide limited interactions with the client throughout the examination. This enables the clinician to observe the client’s ability to manage on his/her own. It is important that the clinician provides a minimum of structure and assistance during the test, even if the client makes an error, because observing their ability to monitor and correct errors without assistance is crucial to showing their independence in consideration of executive functions. The examiner may withhold cueing for up to 10 minutes, unless a situation is judged as dangerous. If the person is clearly unable to perform a component without help the examiner may provide graded assistance (Bottari et al, 2010b).

If the client chooses a task that is not part of the ADL Profile the clinician may ask the client to consider another goal.

What to consider before beginning:

The ADL Profile is best administered in the client’s home and community environment (Bottari et al., 2010b).

Klein et al. (2008) reviewed standardized performance-based ADL measures developed for adult/geriatric populations using an action research study design with 10 occupational therapists working with adult/geriatric clients with physical dysfunction in a tertiary-care rehabilitation hospital, to identify which measures best matched principles of occupational therapy practice and intended outcomes. The ADL Profile achieved the highest rating for its ‘fit’ with the values, beliefs and principles that underpin occupational therapy practice when compared with 17 other ADL measures, including the Assessment of Motor and Process Skills, Rivermead ADL Assessment, the Functional Performance Measure, Nottingham ADL Scale, Barthel Index and Functional Independence Measure. The ADL Profile met four of five construct criteria:

  • Client-centred (score enables item relevancy for client);
  • Dynamic interaction (measure acknowledges dynamic interaction between the client, task and physical environment, but does not consider the financial or social environment);
  • Uniqueness of the individual (measure enables assessment of physical, affective and cognitive performance components); and
  • Uniqueness of performance (measure incorporates client determination of task process unless client safety is a factor).

Of the 18 tools analysed, none achieved a score for the fifth dimension, a holistic perspective (i.e. integration of the client’s roles, culture, resources, spiritual beliefs and values). While it was reported that the ADL Profile did not consider the social environment, it is important to note that the questionnaire is administered to the patient’s significant other.

Scoring and Score Interpretation:

Each task is scored according to independence in task performance (task score) and the manner in which the task is performed (operation score) with regards to the following four operations:

(i)formulate a goal
(ii)plan
(iii)carry out the task
(iv)verify attainment of the initial goal (Bottari et al., 2010b).

Tasks are scored using a four-level ordinal scale:

0 dependent
1 requires verbal assistance (1v) or physical assistance (1p) or verbal and physical assistance (1vp)
2 independent with difficulty
3 independent

Scores are not added across tasks or operations. The task score is determined by the lowest score on any of the four operations observed during performance of the task. Therefore, difficulty in any operation directly influences independence and task performance.

Time:

Time to administer the ADL Profile will depend on the client’s stage of recovery and the number of tasks the clinician needs to administer. In acute care, it may take between 30 and 60 minutes as the clinician may decide to only administer self care tasks and one or two tasks from the community or home domains. When administered in preparation for discharge from a rehabilitation hospital or to community based participants to whom all tasks may be pertinent to administer, up to 7 hours may be required.

The administration time is acceptable when only components of the tool are administered to the subjects, but the assessment is length if administered in full. However, the authors note that the wealth of information obtained from observing the person complete various activities in her home and community environment cannot be underestimated in terms of its contribution to treatment planning.

Training requirements:

The ADL Profile is intended for use by occupational therapists. It is recommended that clinicians complete a three-day training course to ensure correct administration and interpretation. The course provides information regarding the measure (objectives, conceptual frameworks, variables, administration procedure, scoring and interpretation), uses video to provide instruction regarding administration, and provides opportunities to practice task analysis and scoring (Bottari et al., 2010a).

Subscales:

N/A

Equipment:

The ADL Profile does not necessitate specialized equipment but requires any objects the client typically uses in his/her daily living.

Alternative forms of the ADL

There are no other forms of the ADL Profile.

Client suitability

Can be used with:

  • Patients with stroke.
  • Patients with TBI throughout the continuum of care: – to assist in discharge planning from an acute care hospital, in rehabilitation and for community reintegration (Bottari et al., 2006; C. Bottari personal communication, November 6, 2012).
  • Patients with schizophrenia (Semkovska et al., 2004).

Should not be used with:

  • None reported

In what languages is the measure available?

French and English.

Summary

What does the tool measure? The ADL Profile measures independence in everyday activities in consideration of executive function deficits related to goal setting, planning and execution.
What types of clients can the tool be used for? Clients with traumatic brain injury and stroke.
Is this a screening or assessment tool? Assessment tool
Time to administer Time to administer the ADL Profile will depend on the client’s stage of recovery and the number of tasks the clinician needs to administer. In acute care, it may take between 30 and 60 minutes as the clinician may decide to only administer self care tasks and one or two tasks from the community or home domains. When administered in preparation for discharge from a rehabilitation hospital or to community based participants to whom all tasks may be pertinent to administer, up to 7 hours may be required.
Versions ADL Profile
Other Languages French
Measurement Properties
Reliability Internal consistency:
No studies have examined the internal consistency of the ADL Profile when used with an adult stroke population.

Test-retest:
No studies have examined the test-retest reliability of the ADL Profile when used with an adult stroke population.

Intra-rater:
No studies have examined the intra-rater reliability of the ADL Profile when used with an adult stroke population.

Inter-rater:
One study reported adequate inter-rater reliability for three tasks: preparing a hot meal; eating; obtaining information.

Validity Content:
The ADL Profile was established through literature reviews and consultation with expert researchers and clinicians.

Criterion:
No studies have reported on the criterion validity of the ADL Profile when used with an adult stroke population.

Construct:
Convergent:
One study reported significant correlations between five ADL Profile tasks related to personal care and corresponding tasks of the FIM (Standing up, Toilet transfers, Bathtub transfers, Walking, Stair climbing).

Floor/Ceiling Effects No studies have examined ceiling effects of the ADL Profile when used with an adult stroke population.
Sensitivity / Specificity No studies have examined the sensitivity/specificity of the ADL Profile when used with an adult stroke population.
Does the tool detect change in patients? No studies have reported on responsiveness of the ADL Profile when used with an adult stroke population.
Acceptability The administration time is acceptable when only components of the tool are administered, but administration in full may take up to seven hours over several sessions. However, the wealth of information obtained from observing the person complete various activities in her home and community environment cannot be underestimated in terms of its contribution to treatment planning.
Feasibility The ADL Profile can be administered by an occupational therapist. It requires completion of a three-day training course.
How to obtain the tool? Available at the Canadian Association of Occupational Therapists: www.caot.ca or Les Éditions Émersion http://www.leseditionsemersion.com/articles.php?lng=fr&pg=6.

* Initially developed for a traumatic-brain injured population, the psychometric properties of the tool with this population are described in the administration guide of the too

Psychometric Properties

Overview

A literature search was conducted to identify all relevant publications on the psychometric properties of the ADL Profile relevant to individuals with stroke. Two studies were found.

Floor/Ceiling Effects

No studies have reported ceiling effects of the ADL Profile in clients with stroke.

Reliability

Internal consistency:
No studies have reported on the internal consistency of the ADL Profile in clients with stroke.

Test-retest:
No studies have reported on the test-retest reliability of the ADL Profile in patients with stroke.

Intra-rater:
No studies have reported on the intra-rater reliability of the ADL Profile in patients with stroke.

Inter-rater:
Dell’Aniello-Gauthier (1994) reported that the ADL Profile demonstrates adequate inter-rater reliability (mean kappa = 0.58-0.68) for three ADL tasks: preparing a hot meal, eating and obtaining information.

Validity

Content:

The ADL Profile was established through literature reviews & consultation with expert researchers and clinicians (Dutil et al., 2005).

Criterion:

No studies have reported on the criterion validity of the ADL Profile.

Construct:

Convergent:
Gervais (1995) found significant correlations between 5 tasks of the ADL Profile related to personal care and corresponding tasks of the Functional Independence Measure (Kendall’s tau c = 0.40-0.73; p<.001).

Responsiveness

No studies have examined the responsiveness of the ADL Profile.

References

  • Azegami M., Ohira M., Miyoshi K., Kobayashi C., Hongo M., Yanagihashi R., & Sadoyama T. (2007) Effect of single and multi-joint lower extremity muscle strength on the functional capacity and ADL/IADL status in Japanese community-dwelling older adults. Nursing & Health Sciences, 9(3), 168-176.
  • Bottari, C., Dutil, C., Dassa, C., & Rainville, C. (2006) Choosing the most appropriate environment to evaluate independence in everyday activities: Home or clinic? Australian Occupational Therapy Journal, 53, 98-106.
  • Bottari, C., Dassa, C., Rainville, C., & Dutil, C. (2010a). A Generalizability Study of the Instrumental Activities of Daily Living Profile. Archives of Physical Medicine and Rehabilitation, 91, 734-42
  • Bottari, C., Dassa, C., Rainville, C., & Dutil, C. (2010b). The IADL Profile: Development, content validity, intra- and interrater agreement. The Canadian Journal of Occupational Therapy, 77 (2), 90-101.
  • Canadian Association of Occupational Therapists. (2012). ADL Profile. Retrieved from: http://www.caot.ca/default.asp?pageid=1438
  • Crochard, K. (1987). Les activités du GESCOM en 1986. Paris: Centre national d’études des telecommunications.
  • Dell’Anniello-Gauthier, M. (1994). Étude métrologique du mini-profil, instrument de mesure du statut fonctionnel des personnes âgées victimes d’un accident vasculaire cérébral. Sherbrooke, Québec : Université de Sherbrooke.
  • Dutil, E., Bottari, C., Vanier, M., & Gaudreault, C. (2005). ADL Profile: description of the instrument. 4th ed. Montréal: Les Éditions Émersion.
  • Dutil, E., Bottari, C., Vanier, M. & Gaudreault, C. (2005). Profil des AVQ: Description de l’outil, 4th ed. Montréal: Les Éditions Émersion.
  • Fougeyrollas, P. Saint-Michel, G. & Blouin, M. (1989). Propostition d’une révision du 3e niveau de la CIDIH: le handicap. [Proposition for a revision of the 3rd level of the International Classification of Handicaps: the handicap]. Réseau International CIDIH, 2 (1), 9-32.
  • Gervais N. (1995). Comparaison du profil des AVQ et de la mesure d’indépendance fonctionnelle: validité de trait. Montréal: Université de Montréal.
  • Kielhofner, G. (1995). A Model of Human Occupation: Theory and Application. USA: Lippincott Williams & Wilkins.
  • Klein, S., Barlow, I. & Hollis, V. (2008). Evaluating ADL measures from an occupational therapy perspective. Canadian Journal of Occupational Therapy 75,: 69-81.
  • Lawton, P. (1983). Environment and other determinants of well-being in older people. The Gerontologist 23, 349-357.
  • Luria, A.R. (1973). The Working Brain – An Introduction to Neuropsychology. New York: Basic Books.
  • Semkovska, M., Bedard, M.A., Godbout, L., Limoge, F., & Stip, E. (2004) Assessment of executive dysfunction during activities of daily living in schizophrenia. Schizophrenia Research, 69: 289-300

See the measure

How to obtain the ADL Profile:

The ADL Profile can be purchased online at the Canadian Association of Occupational Therapists online store (www.caot.ca) or at Les Éditions Émersion (http://www.leseditionsemersion.com).

Table of contents

Assessment of Motor and Process Skills (AMPS)

Evidence Reviewed as of before: 26-11-2010
Author(s)*: Lisa Zeltzer, MSc OT
Editor(s): Nicol Korner-Bitensky, PhD OT; Elissa Sitcoff, BA BSc
Expert Reviewer: Dianna Robertson, BSc OT, MSc OT (Thesis candidate)

Purpose

The Assessment of Motor and Process Skills (AMPS) is an observational assessment that allows for the simultaneous evaluation of motor and process skills and their effect on the ability of an individual to perform complex or instrumental and personal activities of daily living (ADL). The AMPS is comprised of 16 motor and 20 process skill items.

In-Depth Review

Purpose of the measure

The Assessment of Motor and Process Skills (AMPS) is an observational assessment that allows for the simultaneous evaluation of motor and process skills and their effect on the ability of an individual to perform complex or instrumental and personal activities of daily living (ADL). The AMPS is comprised of 16 motor and 20 process skill items.

Motor skills are the observable goal-directed actions people perform during ADL task performance in order to move themselves or the task objects (e.g. walk, transport objects, reach for and manipulate objects, position the body).

Process skills refer to the ability of an individual to logically sequence the actions of the ADL task performance over time (e.g. initiate and sequence actions, use appropriate tools and material, and accommodate actions when problems are encountered).

The AMPS is based on a ‘top down’ assessment approach. Using a ‘top down’ approach means that the AMPS assessment begins “with the ability of the individual to perform the daily life tasks that he or she wants and needs to perform to be able to fulfill his or her roles competently and with satisfaction” (Fisher, 2003). The ‘top down’ approach involves finding out more about a client’s occupational concerns and then observing the client performing that occupation. Through the observation process, the therapist is able to use clinical reasoning to identify the underlying functional deficit in order to intervene to compensate for the deficit, if this is possible.

The quality of the person’s occupational performance is assessed by rating the effort, efficiency, safety, and independence demonstrated in each of the motor and process skills that comprise the task performance.

Available versions

  • The AMPS was first developed by Fisher in 1990, however, the AMPS was not published by Fisher until 1995. The AMPS manual is currently on its sixth edition.
  • In 2001, it was recommended that 20 new tasks be added to the AMPS (Bray, Fisher, & Duran, 2001). These tasks were added to benefit individuals at the lower or higher ends of the AMPS motor and process skill scales.
  • To date, there are 85 AMPS tasks to select from. A list of these tasks can be found on the AMPS International webpage at: http://www.ampsintl.com/AMPS/resources/tasks.php.

Features of the measure

Items:
There are no actual items to the AMPS. After an initial interview with the caregiver or the client, the rater selects a subset of 3-5 ADL tasks from a list
of standardized tasks that are described in the AMPS manual (e.g. fetching a drink from the fridge, folding laundry, preparing a sandwich). The tasks selected must be relevant and meaningful to the client, and consist of tasks that he/she once knew how to perform. The tasks must be challenging to the client. From this subset of tasks, the client then selects 2-3 tasks to perform.

Prior to beginning task observation, the client and rater must agree on the elements of the task and the tools and materials to be used. In a clinical setting, the client can familiarize his/herself by placing the tools and materials where he/she prefers them to be stored. The client is expected to perform the designated tasks in their usual manner, but must also adhere to the guidelines specified
in the APS manual. For example, if the client selects task A-1, retrieving a beverage from refrigerator, the examiner must watch for the following specific criteria:

  • Obtain container of beverage from the refrigerator
  • Pour the beverage into cup or glass
  • Serve beverage
  • Clean up.
  • Although each of the tasks involves a standard procedure, some
    flexibility is allowed to ensure that the assessment remains
    semi-individualized.

To assist the examiner in preparing for and administering the AMPS interview, task notes are provided which outline the things to look for while task is being completed.

A list of tasks can be found at the AMPS International website: http://www.ampsintl.com/AMPS/resources/tasks.php

One can select online which tasks are to be completed by checking the box beside the task. To find a list of the steps to look for while the task is being completed, select ‘print notes’, which will automatically generate the steps into a printable worksheet. These task notes are intended to be used in combination with the AMPS task descriptions to assist the rater, and are not intended to replace
careful reading of the task descriptions in the manual.

Scoring:
The AMPS uses a 4-point Likert scale to rate the client’s performance on 16 motor and 20 process skills (see table).

Score Interpretation
4 Competent, when the patient performs the task without
evidence of increased effort, decreased efficiency, or lack of
safety.
3 If the examiner questions the effectiveness, the
performance is scored ‘questionable’.
2 Ineffective performance that disrupts or interferes with
the action
1 Marked deficient performance that impedes the action
progression and yields unacceptable outcome.

The raw motor and process scores are entered into the AMPS computer-scoring program and analyzed using many-faceted Rasch analysis (Linacre, 1993). Many-faceted Rasch analysis is used to allow for the calibration of: 1. skills item difficulty, 2. task challenge, 3. individual evaluator leniency, and 4. client variation in ADL ability, on the same linear scale.

The Rasch analysis creates a unique method that enables the AMPS administrator to predict how an individual is expected to perform on any of the calibrated ADL tasks in the assessment after completion of only two or three tasks (Fisher, 1995). This analysis converts the clients’s ordinal raw scores into equal interval measures of ability (person ability measures), which are expressed in log-odds probability units (logits). The logit ability measures are placed on a linear  continuum of increasing ability for each of the ADL scales (motor and process). The AMPS person ability measures represent the person’s place on the continuum, and provides an indication of how challenging a task that person can manage effectively. The higher the ADL motor or ADL process ability, the more able is the client (Fisher, 1997).

If the AMPS is to be used for documenting treatment efficacy, quality assurance, or research, it must be computer scored.

Motor and process cutoff measures:

The position of a person’s ability measures on the ADL motor and ADL process scales can also be evaluated relative to the motor and process cutoff measures. The cutoff measures are 2.0 logits for the ADL motor scale and 1.0 logits for the ADL process scale. These cutoff measures were developed based on the performance of 2,548 subjects in the AMPS database (Fisher, 1997). Individuals with ability measures that fall below the cutoff on either the ADL motor or ADL process scale demonstrated observable motor or process deficits that were
affecting their ability to perform ADL tasks in an effective manner. They were also more likely to require assistance in performing daily life tasks.

Time:
It takes 30-40 minutes to administer the AMPS (AMPS International Website:
http://www.ampsintl.com/AMPS/resources/tasks.php).

Subscales:
The AMPS has two subscales: ADL motor skills and ADL process skills.

Equipment:
No specialized equipment is required to complete the AMPS. Only the AMPS notes, and the relevant equipment for task completion are required.

(AMPS International Website:
http://www.ampsintl.com/AMPS/resources/tasks.php).

Training:
The AMPS can be administered only by occupational therapists who have completed a 5-day training and calibration workshop. Information regarding
training sessions can be found by visiting the AMPS International website: http://www.ampsintl.com/workshops.htm

The AMPS administration manual and computer scoring software is only provided to individuals who participate in AMPS training and calibration workshops.

To become an AMPS Calibrated Rater, an occupational therapy practitioner must complete the following steps:

  • Attend a 5-day training course
  • Test 10 clients who perform 2-3 AMPS tasks
  • Independently interview and score live clients (the use of videotapes is not acceptable). Two of ten clients may be co-scored (two therapists observing a client at the same time, but independently score client performance).
  • Enter the data into the computer using the AMPS computer-scoring program
  • Email exported data to AMPS Project International within 3 months of taking the course.

Alternative forms of the AMPS

  • The School Version of the Assessment of Motor and Process
    Skills (School AMPS).

    The School AMPS is an evaluation tool for measuring student’s schoolwork task performance in typical classroom settings.

Client suitability

Can be used with:

  • Patients with stroke.

Should not be used with:

  • The AMPS cannot be used to diagnose underlying mind-brain-body problems (e.g. memory, apraxia, motivation, perception).
  • The AMPS cannot be administered to patients who are confined to bed or who are unwilling to participate in simple daily living tasks.
  • The AMPS is not suitable for children under the age of 3

Languages of the measure

To date, the AMPS has been administered to over 12,000 subjects from North America, Scandinavia, the United Kingdom, Australia, and New Zealand.

A number of studies have supported the validity of the AMPS as a cross-cultural measure (Fisher, Liu, Velozo & Pan 1992; Goldman & Fisher, 1997; Goto, Fisher & Mayberry, 1996; Magalhaes, Fisher, Bernspang & Linacre, 1996; Stauffer, Fisher & Duran, 2000). For example, Goto, Fisher and Mayberry (1996) tested the cross-cultural validity of the AMPS with six trained raters from diverse backgrounds, and found high cross-cultural validity and inter-rater reliability.

Validation of the AMPS has been established for use in Sweden, (Bernspang & Fisher, 1995), Taiwan (Fisher, Liu, Velozo, & Pan, 1992), and in Spain (http://www.terapia-ocupacional.com/Cursos/Curso_AMPS_Escala_Valoracion_Habilidades_Motoras_Procesamiento_Terapia_Ocupacional.htm).

Limited parts of the AMPS manual(s) and software are available in Japanese, Swedish, Dutch, French, Norwegian, Slovenian, Finnish, and Danish. AMPS International is currently working on new translations in Spanish, Italian, and German.

Summary

What does the tool measure? Motor and process skills and their effect on the ability of an individual to perform complex or instrumental and personal activities of daily living (ADL).
What types of clients can the tool be used for? The AMPS can be used with, but is not limited to patients with stroke.
Is this a screening or assessment tool? Assessment
Time to administer The AMPS takes 30-40 minutes to administer.
Versions The School Version of the Assessment of Motor and Process Skills (School AMPS)
Other Languages Limited parts of the manual(s) and software are available in Japanese, Swedish, Dutch, French, Norwegian, Slovenian, Finnish, and Danish. AMPS International is currently working on new translations in Spanish, Italian, and German.
Measurement Properties
Reliability Internal consistency:
No studies have examined the internal consistency of the AMPS.

Test-rest:
Out of two studies examining the test-rest reliability of the AMPS, both reported excellent test-retest.

Intra-rater:
Only one study has examined the intra-rater reliability of the AMPS and reported excellent intra-rater.

Inter-rater:
No studies have examined the inter-rater reliability of the AMPS.

Validity Criterion:
Concurrent:
Excellent correlations with the Scale of Independent Behavior, the Functional Independence Measure, and the Cambridge Cognitive Examination (CAMCOG) have been reported.

Predictive:
The AMPS score has been found to be predictive of the need for supervision/assistance to live in the community, and home safety for individuals with psychiatric conditions associated with cognitive impairments.

Construct:
Known groups:
AMPS can differentiate between individuals with Multiple Sclerosis and healthy controls; patients with stroke and healthy controls; older adults without disability and people with Alzheimer’s disease who need minimal assistance; people with Alzheimer’s disease who require moderate assistance; individuals with and without psychiatric disorders.

Floor/Ceiling Effects No studies have examined the floor or ceiling effects of the AMPS.
Does the tool detect change in patients? One study examined the responsiveness of the AMPS in a 3-arm drug trial and reported significant differences for instrumental ADL process skills among the 3 conditions, suggesting that the AMPS may be a sensitive measure for detecting change under various study conditions in drug trials.
Acceptability The AMPS cannot be used to diagnose underlying mind-brain-body problems (e.g. memory, apraxia, motivation, perception). The AMPS cannot be administered to patients who are confined to bed or who are unwilling to participate in simple daily living tasks, or for children under the age of 3.
Feasibility The AMPS takes 30-40 minutes to administer, and does not require any specialized equipment. The rater selects a subset of 3-5 ADL tasks (from which the client selects 2-3 to perform) from a list of standardized tasks that are described in the AMPS manual. The AMPS is simple to score and uses a 4-point Likert scale. The scores are then analyzed using an AMPS computer-scoring program. The AMPS can be administered only by occupational therapists who have completed a 5-day training and calibration workshop.
How to obtain the tool? The AMPS manual and software can be purchased online at http://www.ampsintl.com/

Psychometric Properties

Overview

We conducted a literature search to identify all relevant publications on the psychometric properties of the AMPS.

Reliability

Test-retest:
Doble, Fisk, Lewis and Rockwood (1999) examined the test-retest reliability of the AMPS in a sample of 55 elderly adults and reported excellent test-retest coefficients for both the motor and process subscores (r = 0.88 and r = 0.86, respectively).

Fisher (1995) reported that with a sample of older adults (mean age of 80), the test-retest reliability was excellent for both the AMPS motor scale (r = 0.91) and for the AMPS process scale (r = 0.90).

Intra-rater:
Fisher, Liu, Velozo and Pan (1992) reported that in a sample of Taiwanese participants without disability, the AMPS had excellent intra-rater reliability (r = 0.93).

Validity

Criterion:

Concurrent:
Bruininks, Woodcock, Weatherman, and Hill (1985) correlated the AMPS with the Scale of Independent Behavior (Neistadt, 1993) and reported excellent correlations (ranging from r = 0.62 to r = 0.85).

Robinson and Fisher (1996) examined the Functional Independence Measure (Keith, Granger, Hamilton & Sherwin, 1987) (r = 0.62) as well as with the Cambridge Cognitive Examination (CAMCOG), a cognitive component of the Cambridge Mental Disorders of the Elderly Examination (an interview measure of dementia) (Roth et al., 1986) (r = 0.65).

Predictive:
Fisher (1997) reported that 84% of people with ADL motor ability measures below 2.0 logits and 93% of those with ADL process ability measures below 1.0 logits, required supervision or assistance to live in the community. The fact that a higher proportion of people with low ADL process ability measures than with low ADL motor ability measures required some assistance demonstrates that the ADL process scale is a better indicator of need for assistance to live in the community than is the ADL motor scale.

McNulty and Fisher (2001) examined whether the AMPS could predict home safety for individuals with psychiatric conditions associated with cognitive impairments. Moderate positive relationships were found between ADL motor and ADL process ability and home safety in both the clinic and the home. Home ADL process ability was the best predictor of home safety for participants who were categorized as less safe in the study.

Construct:

Known groups:
Doble, Fisk, Fisher, Ritvo and Murray (1994) examined the instrumental ADL performance of 22 community-dwelling patients with mild to moderate Multiple Sclerosis in comparison to participants without disability who were matched for age and gender. Functional competence of the patients with Multiple Sclerosis, as measured by the AMPS, was poorer than that of the control group suggesting that the AMPS can differentiate between individuals with Multiple Sclerosis and healthy controls.

Bernspang and Fisher (1995) administered the AMPS to 71 individuals with right cerebral vascular accident, 76 persons with left cerebral vascular accident, and 83 community-living healthy individuals. Both stroke groups had significantly lower IADL performance than the control participants, suggesting that the AMPS can distinguish between patients with stroke and healthy controls.

Hartman, Fisher, and Duran (1999) administered the AMPS to 329 older adults without disability and 167 people with Alzheimer’s disease who need minimal assistance, and 292 with Alzheimer’s disease who require moderate assistance. In this study, the AMPS was able to distinguish between the three groups.

Pan and Fisher (1994) examined the hypothesis that mean AMPS scores would differ between individuals with psychiatric disorders and individuals without. Sixty participants, 30 without and 30 with psychiatric disorders, were studied. The hypothesis was supported for both the AMPS motor and process scales, suggesting that the AMPS can distinguish between individuals with and without psychiatric disorders.

Responsiveness

One pharmacological pilot study of individuals with Alzheimer’s disease examined the responsiveness of the AMPS using repeated measures ANOVA (Oakley & Sunderland, 1997). Significant differences were found for instrumental ADL process skills, but not for motor skills, among three drug conditions. The results of this study suggest that the AMPS may be a sensitive measure for detecting change under various study conditions in drug trials.

References

  • Bernspang, B., Fisher, A. (1995). Differences between persons with right or left cerebral vascular accident on the Assessment of Motor and Process. Archives of Physical Medicine and Rehabilitation, 76, 1144-1151.
  • Bray, K., Fisher, A. G., Duran, L.(2001).The validity of adding new tasks to the Assessment of Motor and Process Skills. American Journal of Occupational Therapy 55,, 409-415.
  • Bruininks, R. H., Woodcock, R. W., Weatherman, R. F., Hill, B. K. (1985). Development and Standardization of the Scales of Independent Behavior. Allen, TX: DLM Resources.
  • Cooke, K, Z., Fisher, A. G., Mayberry, W., Oakley, E. (2000). Differences in activities of daily living process skills of persons with and without Alzheimer’s disease. Occupational Therapy Journal of Research, 20, 87-104.
  • Dickerson, A. E., Fisher, A. G. (2000). Age differences in functional performance. American Journal of Occupational Therapy, 47, 686-692.
  • Doble, S. E., Fisk, J. D., Fisher, A., Ritvo, P. Murray, T. (1994). Functional competence of community-dwelling persons with multiple sclerosis using the Assessment of Motor and Process Skills. Archives of Physical Medicine and Rehabilitation, 75, 843-851.
  • Doble, S. E., Fisk, J. D., Lewis, N., Rockwood, K. (1999). Test-retest reliability of the Assessment of Motor and Process Skills. Occupational Therapy Journal of Research, 19, 203-215.
  • Doble, S. E., Fisher, A. G., Fisk, J. D., MacPherson, K. M. (1992). Validation of the Assessment of Motor and Process Skills (AMPS) with Elderly Adults with Dementia. Final Report to the Alzheimer’s Association. Halifax, Nova Scotia: Dalhousie University.
  • Duran, L., Fisher, A. (1996). Male and female performance on the Assessment of Motor and Process Skills. Archives of Physical Medicine and Rehabilitation, 77, 1019-1024.
  • Fisher, A. G. (1990). Assessment of Motor and Process Skills. Research edition, R. Unpublished test manual. Chicago, IL: University of Illinois at Chicago.
  • Fisher, A. (1995). The Assessment of Motor and Process Skills (AMPS). Fort Collins, CO: Three Star Press.
  • Fisher, A. G. (1997). Assessment of Motor and Process skills, 2nd edn. Fort Collins, CO: Three Star Press.
  • Fisher, A. G., Liu, Y., Velozo, C., Pan, A. W. (1992). Cross-cultural assessment of process skills. American Journal of Occupational Therapy, 46, 876-885.
  • Fisher, A. G. (2003). AMPS: Assessment of Motor and Process Skills. Volume 1: Development, Standardisation, and Administration Manual. 5th edn. Colorado: Three Star Press Inc.
  • Goldman, S., Fisher, A. G. (1997). Cross-cultural validation of the Assessment of Motor and Process Skills (AMPS). British Journal of Occupational Therapy, 46, 77-85.
  • Goto, S., Fisher, A. G., Mayberry, W. L. (1996). Assessment of Motor and Process Skills applied cross-culturally to the Japanese. American Journal of Occupational Therapy, 50, 798-806.
  • Hartman, M. L., Fisher, A. G., Duran, L. (1999). Assessments of functional ability of people with Alzheimer’s disease. Scandinavian Journal of Occupational Therapy, 6, 111-118.
  • Keith, R., Granger, C., Hamilton, B., Sherwin, F. (1987). The Functional Independence Measure: A new tool for rehabilitation. In: N. Eisenberg & R. Grzesiak (Eds.), Advances in Clinical Rehabilitation. New York: Springer.
  • Linacre, J. M. (1993). Many-Facet Rasch Measurement, 2nd edn. Chicago: MESA.
  • Linden, A., Boschian, K., Eker, C., Schalen, W., Nordstrom, C.-H. (2005). Assessment of motor and process skills reflects brain-injured patients ability to resume independent living better than neuropsychological tests. Acta Neurol Scand, 111, 48-53.
  • Magalhaes, L., Fisher, A., Bernspang, B., Linacre, J. (1996). Cross-cultural assessment of functional ability. The Occupational Therapy Journal of Research, 16(1), 45-63.
  • McNulty, M. C., Fisher, A. G. (2001). Validity of using the Assessment of Motor and Process Skills to estimate overall home safety in persons with psychiatric conditions. Am J Occup Ther, 55(6), 649-655.
  • Neistadt, M. E. (1993). A meal preparation treatment protocol for adults with brain injury. Am J Occup Ther, 48, 431-438.
  • Oakley, F., Sunderland, T. (1997). Assessment of Motor and Process Skills as a measure of IADL functioning in pharmacologic studies of people with Alzheimer’s disease: A pilot study. International Psychogeriatrics, 9, 197-206.
  • Pan, A. W., Fisher, A. G. (1994). The Assessment of Motor and Process Skills of persons with psychiatric disorders. American Journal of Occupational Therapy, 48, 775-780.
  • Robinson, S., Fisher, A. G. (1996). A study to examine the relationship of the Assessment of Motor and Process Skills (AMPS) to other tests of cognition and function. British Journal of Occupational Therapy, 59, 260-63.
  • Roth, M., Mountjoy, C., Huppert, F., Hendrie, H., Verna, S., Godard, R. (1986). CAMDEX. The Cambridge Examination for Mental Disorders of the Elderly. Cambridge, UK: Cambridge University Press.
  • Stauffer, L. M., Fisher, A. G., Duran, L. (2000). ADL Performance of black Americans and white Americans on the Assessment of Motor and Process Skills. American Journal of Occupational Therapy, 54, 607-613.

See the measure

How to obtain the AMPS

The AMPS manual and software can be purchased online at http://www.ampsintl.com/

Table of contents

Barthel Index (BI)

Evidence Reviewed as of before: 07-10-2015
Author(s)*: Katie Marvin, PT; Lisa Zeltzer, MSc OT
Editor(s): Annabel McDermott, OT; Nicol Korner-Bitensky, PhD OT; Elissa Sitcoff, BA BSc

Purpose

The Barthel Index (BI) measures the extent to which somebody can function independently and has mobility in their activities of daily living (ADL) i.e. feeding, bathing, grooming, dressing, bowel control, bladder control, toileting, chair transfer, ambulation and stair climbing. The index also indicates the need for assistance in care. The BI is a widely used measure of functional disability. The index was developed for use in rehabilitation patients with stroke and other neuromuscular or musculoskeletal disorders, but may also be used for oncology patients.

In-Depth Review

Purpose of the measure

The Barthel Index (BI) measures the extent to which somebody can function independently and has mobility in their activities of daily living (ADL) i.e. feeding, bathing, grooming, dressing, bowel control, bladder control, toileting, chair transfer, ambulation and stair climbing. The index also indicates the need for assistance in care.

The BI is a widely used measure of functional disability. The index was developed for use in rehabilitation patients with stroke and other neuromuscular or musculoskeletal disorders, but may also be used for oncology patients.

Available versions

The BI was first developed by Mahoney and Barthel in 1965 and later modified by Collin, Wade, Davies, and Horne in 1988.

  • Original 10-item version (Mahoney & Barthel, 1965). Refers to the following 10 categories: feeding, bathing, grooming, dressing, bowel control, bladder control, toileting, chair transfer, ambulation and stair climbing. Items are weighted according to the level of nursing care required and are rated in terms of whether individuals can perform activities independently, with some assistance, or are dependent (scored as 10, 5 or 0).

Features of the measure

Items:

The original 10-item form of the BI consists of 10 common ADL activities including: feeding, bathing, grooming, dressing, bowel control, bladder control, toileting, chair transfer, ambulation and stair climbing. Items are rated in terms of whether individuals can perform activities independently, with some assistance, or are dependent (scored as 10, 5 or 0). Items are weighted according to the level of nursing care required.

Scoring:

The score of the BI is a summed aggregate and there is preferential weighting on mobility and continence. The scores are allotted in the following way: 0 or 5 points per item for bathing and grooming; 0, 5, or 10 points per item for feeding, dressing, bowel control, bladder control, toilet use, and stairs; 0, 5, 10, or 15 points per item for transfers and mobility. The Index yields a total score out of 100 – the higher the score, the greater the degree of functional independence (McDowell & Newell, 1996). This score is calculated by simply totaling the individual item scores, which requires simple arithmetic computation
by hand.

A modified scoring system has been suggested by Shah, Vanclay, & Cooper (1989) using a 5-level ordinal scale for each item to improve sensitivity to detecting change (1=unable to perform task, 2=attempts task but unsafe, 3=moderate help required, 4=minimal help required, 5=fully independent). Shah and coll. (1989) note that a score of 0-20 suggests total dependence, 21-60 severe dependence, 61-90 moderate dependence and 91-99 slight dependence.

Subscales:

None typically reported.

Equipment:

To administer the BI, one only needs a pencil and the test items.

Training:

Administration of the BI does not require training and has been shown to be equally reliable when administered by skilled and unskilled individuals (Collin & Wade, 1988). The BI can also be self-administered (McGinnis, Seward, DeJong, & Osberg, 1986). However, for patients older than 75 years of age, it is not recommended that the BI be administered as a self-report measure (Sinoff & Ore, 1997). One study suggests that the scale can be administered reliably over the telephone (Korner-Bitensky & Wood-Dauphinee, 1995).

Time:

The BI can take as little as 2-5 minutes to complete by self-report and up to 20 minutes to complete by direct observation (Finch, Brooks, Stratford, & Mayo, 2002).

Alternative forms of the BI

  • Modified 10-item version (MBI)(Collin et coll., 1988). Functional categories may be scored from 0 to 1, 0 to 2, or 0 to 3, depending on the item. Total scores range from 0 to 20.
  • 5-item short form(Hobart & Thompson, 2001). The 5-item version refers to the following 5 categories: transfers, bathing, toilet use, stairs, and mobility. Each item is scored 0 to 1, 0 to 2, or 0 to 3, depending on the function. Total scores range from 0 to 20. Hobart & Thompson (2001) found that the 5-item BI is psychometrically equivalent to the 10-item BI (correlation with original version was r = 0.90).
  • The expanded 15-item version(Granger et coll., 1979; Fortinsky & Granger, 1981). Added a 4-point scale of intact/limited/helper required/null. Scores range from 0 to 100. In the 15-item version, a score of 60 is commonly considered to be the threshold score for marked dependence (Granger, Sherwood, & Greer, 1977). High correlations of the expanded 15-item BI and other measures of function have been demonstrated (e.g., with Katz Indice of Activities of Daily Living, r = 0.78; with PULSES profile (medical status, upper and lower limb function, sensory and excretory function, mental and emotional status), r = -0.74 to -0.90 (Shinar, Gross, Bronstein, Licara-Gehr, Eden, Cabrera, et coll., 1987; Granger, 1985; Rockwood, Stolee & Fox, 1993). Scores were also predictive of return to independent living after 6 months (Granger, Hamilton, Gresham, & Kramer, 1989).
  • The extended BI (EBI)(Prosiegel, Bottger, & Schenk, 1996). The EBI consists of 16 items, 15 of which are identical to the Functional Independence Measure. Very little literature exists on the EBI, however Jansa, Pogacnik, and Gompertz (2004) found it to be a reliable and valid measure of disability/activity levels in 33 patients with newly diagnosed acute ischemic stroke.
  • The 3-item BI(Ellul, Watkins, & Barer, 1988).Based on 3 items (bed-chair transfers, mobility, and bladder incontinence), it is a useful alternative to the full BI for assessing function at hospital discharge. To date, this version has only been validated in patients with stroke.
  • Self-rating BI(SB). The SB has good concurrent validity and is well related with the original BI and the Functional Independence Measure. The indexes test-retest reliability is sufficiently high for practical use (Hachisuka, Ogata, Ohkuma, Tanaka, & Dozono, 1997; Hachisuka, Okazaki, & Ogata, 1997; McGinnis et coll., 1986).
  • Early Rehabilitation Barthel Indice (ERI). An extension of the BI, it was developed to assess functioning of individuals with severe brain damage, who often cannot be differentiated appropriately due to floor effects that occur with increasing severity of neurological impairment. The ERI looks at the following aspects: state requiring temporary intensive medical monitoring, tracheostoma requiring special treatment (suctioning), intermittent artificial respiration, confusional state requiring special care, behavioural disturbances requiring special care, swallowing disorders requiring special care, and severe communication deficits. Schonle (1995) found that the ERI is quick, economical, and reliable when administered to 210 early rehabilitation patients and 312 patients with severe brain damage.

There is little consensus over which should be considered the definitive version of the BI (McDowell & Newell, 1996), but the original and the 10-item and 15-item modifications are the most commonly used.

Client suitability

Can be used with:

  • Patients with stroke.

The BI is a frequently used stroke outcome measure. It has been repeatedly shown to be a reliable and valid measure of basic Activities of Daily Living (Mahoney & Barthel, 1965; Loewen & Anderson, 1990; Gresham, Phillips & Labi, 1980; Collin et coll., 1988; Roy, Tongeri, Hay, & Pentland, 1988; Wade & Hewer, 1987; Leung et coll., 2007). In patients with stroke, the BI determines the extent of post-stroke disability, self-care activities and ability to live independently. The total score of the BI has also been found to predict length of stay in hospital (Granger, Albrecht, & Hamilton, 1979).

There are no prerequisites for completing the BI. For patients who are unable to respond to the BI independently, the BI can be completed by proxy (eg. Duncan, Lai, Tyler, Perera, Reker, & Studenski, 2002; Wyller, Sveen, & Bautz-Holter 1995). Further, the BI can be reliably administered over the telephone to either the patient or their proxy (Korner-Bitensky & Wood-Dauphinee, 1995).

Should not be used in:

  • To capture significant losses in higher levels of physical function or activities that are necessary for independence in the home and community. This means that patients can still score a maximum score of 100 and experience significant impairments (Kelly-Hayes et al., 1998).
  • It should be used with caution in patients with mild stroke. It is responsive to change but has definite ceiling effects in persons with mild stroke (Wade & Hewer, 1987; Skilbeck, Wade, Hewer, & Wood, 1983).

In what languages is the measure available?

The BI has been translated and validated in:

  • Dutch (Post, van Asbeck, van Dijk, & Schrijvers, 1995)
  • German (Heuschmann et al., 2005; Valach, Signer, Hartmeier, Hofer, & Steck, 2003)
  • Turkish (Kucukdeveci, Yavuzer, Tennant, Suldur, Sonel, & Arasil, 2000)
  • Persian (Oveisgharan, 2006)
  • French (Condouret et al., 1988; Wirotius & Foucher-Berres, 1991)
  • Chinese (Leung, Cha, & Shah, 2007) (modified Barthel Index)

Summary

What does the tool measure? Activities of Daily Living
What types of clients can the tool be used for? Patients with stroke, patients with other neuromuscular or
musculoskeletal disorders, oncology patients
Is this a screening or assessment tool? Assessment
Time to administer Self report: 2-5 minutes; Direct observation: 20 minutes,
but may vary according to patient’s abilities and tolerance
Versions Modified 10-item version (MBI); 5-item short form; The expanded 15-item version; The extended BI (EBI); The 3-item BI; Self-rating BI (SB); Early Rehabilitation Barthel Index (ERI)
Other Languages Dutch, German, Turkish, Persian, French, Chinese
Measurement Properties
Reliability Internal consistency:
Five studies of the MBI reported excellent internal consistency.

Test-retest:
One study of the MBI reported excellent test-retest reliability.

Inter-rater:
One study of the MBI and four studies of BI reported excellent inter-rater reliability; and one study of the BI reported adequate inter-rater reliability.

Validity Criterion:
Concurrent:
One study demonstrated excellent concurrent validity between the MBI and motor-Functional Independence Measure (FIM) at admission and discharge.

Predictive:
The MBI predicted instrumental ADL permformance at 6-months post-stroke; likelihood a patient will regain continence following stroke; risk for falls in patients with stroke; functional recovery following stroke; and acute care hospital length following stroke.

Construct:
Excellent correlations in patients with stroke on the physical mobility dimension of the Nottingham Health Profile Subscale; the Physical Functioning subscale of the SF-36; Berg Balance Scale; the Fugl-Meyer Assessment Scale; Frenchay Activities Indice.

Does the tool detect change in patients?
  • Significant ceiling effects noted for the BI, meaning that it doesn’t detect change well in highly functional individuals.
  • The Functional Independence Measure was developed as a measure that would be better able to detect change in disability than the BI, however little to no difference has been found.
  • Out of 8 studies examined, 3 reported that the BI had a large ability to detect change, 3 reported adequate ; 2 reported small.
Acceptability The MBI/BI has been evaluated for both self-report and use with proxy respondents in addition to direct observation.
Feasibility The MBI/BI is simple to administer. Requires training if administered by direct observation. It has been developed in many forms that can be administered in many situations and can be used for longitudinal assessment.
How to obtain the tool? For a copy of the original BI click here http://www.strokecenter.org/trials/scales/barthel.pdf

Psychometric Properties

Overview

There is considerable psychometric data available for the BI (McDowell & Newell, 1996) and its various modified versions. For the purposes of this review, we conducted a literature search to identify all relevant publications on the psychometric properties of the original BI and the modified 10-item BI (MBI), the two most commonly used versions. We then selected to review articles from high impact journals, and from a variety of authors.

*Please note that the content in the original BI and MBI version of the BI is the same. Only the scoring values were changed in the MBI version (scored 0, 1, 2 or 3 versus 0, 5 and 10 in the original version), and thus do not impact the clinimetric properties of the tool (Quinn, Langhorne and Stott, 2011). The MBI yields a score ranging from 0 to 20, whereas the original BI yields a score of 0 to 100. For the purposes of this module, the psychometric properties for both the BI and MBI will be presented together and will be referred to as either the BI or MBI.

Floor and ceiling effect

Salbach et coll. (2001) examined the ceiling effects of the BI, Timed Up and Go (TUG), Berg Balance Scale (BBS), 10 meter walk test (10mWT) and 5 meter walk test (5mWT) in 50 patients with residual gait deficits after a first-time stroke. The BI demonstrated the most significant ceiling effects at both 8 and 38 days post-stroke (28% and 56% respectively).

Dromerick, Edwards and Diringer (2003) examined the floor/ceiling effects of the BI, the Functional Independence Measure (FIM), the Modified Rankin Scale (MRS) and the International Stroke Trial Measure. The four measures were administered to 95 patients with stroke on admission to and discharge from rehabilitation. The BI demonstrated adequate floor effects at admission (5%) and poor ceiling effects at discharge (27%), whereas the FIM demonstrated excellent floor and ceiling effects (0% for both); the MRS demonstrated adequate floor effects at admission (18%) and excellent ceiling effects at discharge (0%); and the International Stroke Trial Measure demonstrated poor floor effect at admission (100%) and excellent ceiling effect at discharge (0%).

Van der Putten, Hobart, Freeman and Thompson (1999) compared the floor/ceiling effects of the MBI to that of the Motor-FIM, cognitive-FIM and total FIM in 201 patients with multiple sclerosis and 82 patients with stroke undergoing inpatient neurorehabilitation. The MBI, and motor-FIM demonstrated adequate floor and ceiling effects for both patients with stroke and patients with multiple sclerosis (floor effects = 1.2% (BI, stroke), 1.2% (motor-FIM, stroke); and ceiling effects = 8.5% (BI, stroke) and 1.2% (motor-FIM, stroke). The total-FIM showed no floor or ceiling effects for both patients with stroke and patients with MS (0% for all). The Cognitive-FIM demonstrated poor ceiling effects in patients with multiple sclerosis (36%) and adequate ceiling effects in patients with stroke.

Hsueh, Lin, Jeng and Hsieh (2002) compared the floor/ceiling effects of the FIM to that of the MBI and the 5-item BI (BI-5) in 118 patients with stroke undergoing treatment on an inpatient rehabilitation unit. The MBI and the motor-FIM both exhibited adequate floor effects at admission and discharge (MBI 18.2% and 4.7%; motor-FIM 5.8% and 3.5% respectively) and excellent ceiling effects at admission and discharge (0% for all). The BI-5 exhibited poor floor effects at admission (46.6%) and adequate floor effects at discharge (13.6%), and excellent ceiling effects at admission and discharge. The results of this study indicate that the MBI and motor-FIM have comparable floor/ceiling effects, with the motor-FIM performing slightly better with respect to floor effects (18.2% vs. 5.8%).

Reliability

Internal consistency:
Hobart and Thompson (2001) compared the psychometrics of the MBI, FIM and the 30-item FIM + Functional Assessment Measure (FIM+FAM) in 149 patients with various neurological disorders. All measures were found to be psychometrically similar measures of physical disability. The internal consistency of the MBI was excellent, with a Cronbach’s alpha of 0.94 (Cronbach’s alpha of the FIM ranged from 0.89-0.96).

Hsueh, Lin, Jeng and Hsieh (2002) compared the internal consistency of the FIM to that of the MBI and the 5-item BI (BI-5) in 118 patients with stroke undergoing treatment on an inpatient rehabilitation unit. The MBI and FIM motor subscale both demonstrated excellent internal consistency (Cronbach’s alpha coefficient ≥ 0.84), whereas the BI-5 demonstrated adequate internal consistency (Cronbach’s alpha coefficient ≥ 0.71) at admission and discharge.

Quinn, Langhorne and Stott (2011) conducted a literature review examining the internal consistency of the MBI in studies involving patients with stroke. The internal consistency of the MBI was found to be excellent (Cronbach’s alpha ³ 0.80) across all reviewed studies, as detailed below.

Shah and coll. (1989) examined the internal consistency of the MBI in 258 patients with stroke. The internal consistency was excellent (Cronbach’s alpha 0.90).

Leung and coll. (2007) examined the internal consistency of the Chinese version and the English version of the MBI and found internal consistency to be excellent for both measures (Cronbach’s alpha 0.93 and 0.92 respectively).

Hseuh, Lee and Hsieh (2001) examined the internal consistency of the MBI in 121 Taiwanese patients with stroke at four time points (14, 30, 90 and 180 days post-stroke). The internal consistency of the BI was excellent (Cronbach’s alpha 0.89-0.92).

Test-retest:
Green, Forster and Young (2001) examined the test-retest reliability of the MBI, Rivermead Mobility Indice (RMI), Nottingham extended Activities of Daily Living Scale (NEADL) and Frenchay Activities Indice (FAI) in 22 patients that were at least one year post-stroke. The four measures were administered twice, with a one-week interval. The MBI and RMI were found to have the strongest test-retest reliability with 75% and 85% agreement overall, respectively; however there was still considerable variability in kappa statistics (BI kappa =-0.09-0.81; RMI kappa =0.64-1.00). The NEADL and FAI demonstrated greater variability and more error (NEADL kappa =0.14-0.89; FAI kappa =0.25-1.00).

Inter-rater:
Leung and coll. (2007) examined the inter-rater reliability of the Chinese and English versions of the MBI in 15 patients with stroke. The inter-rater reliability was found to be excellent for the Chinese version (kappa = 0.81-1.00) and adequate to excellent for the English version (kappa =0.63-0.85), as calculated using kappa statistics.

Duffy, Gajree, Langhorne, Stott and Quinn (2013) conducted a systematic review examining the inter-rater reliability of the BI and MBI in patients with stroke. In a systematic review and meta-analysis, 10 studies were included that involved assessors of differing backgrounds and experience. The BI was found to have excellent inter-rater reliability in eight of the ten studies and adequate inter-rater reliability in two of the ten studies, as calculated using intraclass correlation (ICC), kappa statistics or weighted kappa statistics (ICC ranging from 0.94 to 0.96; kappa ranging from 0.62 to 0.90; weighted kappa ranging from 0.70 to 0.99). The results from five of the 10 studies are included below; the remaining 5 studies could not be reviewed for the purposes of this module as they were not available in English.

Loewen and Anderson (1988) examined the inter-rater and intra-rater reliability of the BI in seven patients with stroke. Inter-rater reliability and intra-rater reliability were excellent (ICC=0.96 and 0.99 respectively).

Wolfe, Taub, Woodrow and Burney (1991) compared the inter-rater and intra-rater reliability of the BI with the Rankin Scale. Inter-rater reliability was excellent for both the BI and Rankin Scale (kw=0.88 to 0.98 and 0.75 to 0.95 respectively). Intra-rater reliability was excellent for both the BI and Rankin Scale (kw=0.98 and 0.95 respectively).

Hseuh, Lee and Hsieh (2001) examined the inter-rater reliability of the BI in Taiwanese patients with stroke, at four time points (14, 30, 90 and 180 days post-stroke). The inter-rater reliability between items of the BI was adequate (weighted kappa = 0.53) to excellent (weighted kappa =0.94). The inter-rater reliability for the total score was excellent (ICC=0.94).

Oveisgharan and coll. (2006) examined inter-rater reliability of a Persian translated version of the BI; inter-rater reliability was excellent (weighted kappa =0.99).

Cincura and coll. (2008) examined the inter-rater reliability of the National Institutes of Health Stroke Scale, Modified Rankin Scale and the BI in Brazilian patients with stroke. Inter-rater reliability was found to be adequate (kappa =0.70).

Validity

Content:

No studies have examined the content validity of the BI in patients with stroke.

Criterion:

Concurrent:
Hsueh, Lin, Jeng and Hsieh (2002) examined the concurrent validity of MBI and the 5-item BI (BI-5) with the motor subscale of the FIM in patients with stroke, using Spearman correlation coefficient. The three measures were administered to 118 patients with stroke at admission to and discharge from an inpatient rehabilitation unit. Concurrent validity of the MBI and the FIM motor subscale was excellent at admission and discharge (r=0.92 and 0.94 respectively), whereas the 5-item BI demonstrated adequate to excellent concurrent validity with the FIM motor subscale at admission and discharge (r=0.74 and 0.92 respectively).

Predictive:
Hseuh, Lee and Hsieh (2001) examined the predictive validity of the MBI in 121 patients with stroke by comparison with the Frenchay Activities Indice (FAI), using Pearson product-moment correlation coefficient. The MBI was administered at 14, 30, 90 and 180 days post-stroke and the FAI was administered at 180 days post-stroke. The MBI scores at 14, 30 and 90 days post-stroke demonstrated adequate correlation with FAI scores at 180 days post-stroke, (r=0.59, 0.66. 0.63 respectively). Results of this study found the MBI to be an adequate predictor of instrumental ADL performance at six months following stroke onset.

Patel, Coshall, Lawrence, Rudd and Wolfe (2001) examined the ability of the MBI and Frenchay Activity Indice (FAI) to predict whether a patient with post-stroke urinary incontinence would regain continence. The study involved 207 patients with stroke with new onset urinary incontinence in the acute phase of recovery. Univariate analysis and multiple regression analysis were used to determine predictive validity. The MBI and the FAI were administered on approximately day seven post-stroke to allow for medical stabilization and at 3-months post-stroke. Patients scoring 15 to 18 (out of 20) on the MBI on day seven were found more likely to regain continence as compared with those scoring less than 15 (Odds ratio=21.8, 95% CI=5.95 – 79.7). At 3 months, patients with incontinence were found to have greater disability as measured by the MBI (P<0.001) and FAI (P=0.002) and greater rates of institutionalization (P<0.001).

Sze, Wong, Leung and Woo (2001) investigated the predictors of falls in patients with stroke, using a study sample of 677 patients admitted to an inpatient rehabilitation stroke unit. Initial assessments, including the MBI, were completed on admission (three to seven days following stroke onset). For the purposes of their study, MBI scores were stratified as: ≥15 mild disability, 6-14 moderate to severe disability, and ≤5 very severe disability. Patients with moderate to severe disability (MBI scores 6-14) were found to have an increase risk for falls (odds ration 2.59, 95%CI=1.24-5.42, r=0.0114). Dysphagia was also found to put patients at an increased risk for falls (odds ratio 1.81; 95% CI, 1.03–3.17, r=.0382).

Tilling and coll. (2001) examined the ability of the MBI to predict functional recovery following stroke. The MBI was administered to 299 patients with stroke at baseline, 2, 4, 6, and 12 months following stroke; recovery trajectories were then plotted using the MBI scores in an effort to establish a prediction model based on the found normal patterns of recovery. Performance of the prediction model was validated using an additional group of 710 patients with stroke. Initial MBI scores, when considered along with individual patient characteristics (such as age, sex and pre-stroke disability), were found to be predictive of future MBI scores up to 1-year following stroke. The predictive validity was found to be even stronger when the patient’s actual observed recovery was taken into consideration and the predictions of future MBI scores were adjusted accordingly. Scoring <1 point below the predicted score on the MBI was found to be predictive of death before the next assessment time point (65% sensitivity, 79% specificity). The results of this study suggest that this model can aid in establishing initial recovery predictions, developing rehabilitation goals and monitoring recovery in patients with stroke.

Chang, Tseng, Weng, Lin, Liou and Tan (2002) examined the predictors of acute care hospital length of stay in 330 patients with first-ever acute stroke. Univariate analysis and multiple regression analysis were used to determine predictive validity. MBI scores at admission (r=0.042), along with National Institute of Health Stroke Scale (NIHSS) scores at admission (r=0.001), the quadratic term of initial NIHSS score (r=0.001), small-vessel occlusion stroke (r<0.001), gender (male) (r=0.004) and smoking (r=0.043) were found to be the most significant predictors of hospital length of stay. A one-point decrease in score on the MBI (indicating a decline in function) corresponded to an increase in length of stay by approximately one day.

Hsieh and coll. (2007) investigated the minimal clinically important difference (MCID) of the modified 10-item BI in a two-part study involving patients with sub-acute to chronic stroke. In the initial part of the study, 43 patients with sub-acute stroke that demonstrated potential for improvement with regard to activities of daily living (ADL) were selected for a 4-week intensive occupational therapy program. The MBI and a 15-point Likert-type scale assessing the patients’ perceived global ratings of their ADL function were administered at baseline and at discharge (with a mean interval between assessment and discharge of 25 days). The estimated MCID was 1.85. The second part of the study involved assessing the repeatability of scores in 56 patients with chronic stroke who were thought to have stable ADL function. The estimated MCID was 1.45. Results indicate that an improvement in total score by 1.85 points or more (on the 0 to 20 scoring scale) indicate a meaningful change beyond measurement error, and thus a change in score less than 1.85 points may be subject to measurement error.
Note: The MCID estimated in this study is applicable only for improvement in function, not deterioration.

Construct:

Wilkinson and coll. (1997) investigated the construct validity of the MBI as a standard long-term outcome measure of patients with stroke. The Hospital Anxiety and Depression Scale (HADS), London Handicap Scale (LHS), Frenchay Activities Indice (FAI), SF36, Nottingham Health Profile (NHP) and the Life Satisfaction Indice (LSI) were administered alongside the MBI in a long-term study involving 106 patients with first-ever stroke (patients were followed for a mean interval of 4.9 years). Rank Correlation Coefficients were excellent between the MBI and SF36 Physical Functioning dimension (r=0.81), NHP Energy (r=0.605) and Physical Mobility (r=0.840) dimensions, LHS (r=0.726) and FAI (r=0.826). Rank Correlation Coefficients were adequate between the MBI and the SF36 Social Functioning (r=0.481), Role: Physical (r=0.415), Mental Health (r=0.332), Vitality (r=0.500), Bodily Pain (r=0.356) and General Health (r=0.438) dimensions, HADS (r=-0.563), and LSI (r=0.361). Poor correlations were found between the MBI and the SF36 Role: Emotional dimension (r=0.217) and NHP Sleep dimension (r=0.189). The results of this study suggest that the MBI should be administered alongside other measures that assess the psychosocial dimensions of health status as the MBI fails to sufficiently assess these aspects.

Convergent/Discriminant:
Hseuh, Lee and Hsieh (2001) examined the convergent validity of the MBI, Berg Balance Scale (BBS) and the Fugl-Meyer Motor Assessment (FMA) in 121 patients with stroke, using Pearson product-moment correlation coefficient . The three measures were administered at 14, 30, 90 and 180 post-stroke. The total MBI score had excellent correlation with the FMA and BBS scores at all four time points (MBI and FMA r=0.8, 0.81, 0.78, 0.8; MBI and BBS r =0.89, 0.94, 0.9, 0.91 respectively).

Known Groups:
No studies have examined the known groups validity of the BI in patients with stroke.

Responsiveness

Wood-Dauphinee, Williams and Shapior (1990) compared the responsiveness of the BI to the Fugl-Meyer Assessment (FMA) in 167 patients with stroke. Patients were assessed at admission to hospital and at 5-weeks post-stroke. The correlation between mean change in FMA Upper and Lower Extremity Motor subscores and total Barthel Indice scores was adequate (r = 0.57), as calculated using Pearson correlation coefficients. The FMA and BI were both found to have small effect sizes (ES = 0.24 and 0.42 respectively) from admission to 5-weeks post stroke. The results of this study suggest that both measures have poor responsiveness with the BI being more sensitive to detecting change that the FMA.

Salbach and coll. (2001) examined the responsiveness of the BI, Timed Up and Go (TUG), Berg Balance Scale (BBS), 10 meter walk test (10mWT) and 5 meter walk test (5mWT) in 50 patients with residualgait deficits after a first-time stroke. The BI, BBS 5mWT and 10mWT demonstrated large effect sizes and the TUG demonstrated a moderate effect size, between 8 days and 38 days post-stroke, as calculated using standardized response means (SRM = 0.99, 1.04, 1.22, 0.92 and 0.73 respectively).

Hsueh, Lin, Jeng, and Hsieh (2002) compared the responsiveness of the BI, 5-item short form BI (BI-5) and motor-FIM in 118 patients with stroke undergoing treatment on an inpatient rehabilitation unit. The BI, BI-5 and motor-FIM all exhibited high responsiveness, as calculated using standardized response mean (SRM) (BI=1.2; 5-BI=1.2; motor-FIM=1.3) indicating significant sensitivity for detecting change.

Wallace, Duncan, and Lai (2002) compared the responsiveness of the BI to that of the motor-FIM for recovery following stroke. Change was measured using the Modified Rankin Scale. The BI and motor-FIM were administered to 372 patients with stroke at one and three months following stroke. The BI and motor-FIM were both found to have small effect sizes (ES = 0.31 and 0.28 respectively), indicating similar responsiveness between the measures.

Van der Putten, Hobart, Freeman and Thompson (1999) compared the responsiveness of the MBI to that of the motor and cognitive components of the FIM and the FIM total score in 201 patients with multiple sclerosis and 82 patients with stroke undergoing inpatient neuro-rehabilitation. The MBI and the total-FIM and motor-FIM all demonstrated large effect sizes for patients with stroke (ES = 0.95, 82, 91 respectively) and the cognitive-FIM demonstrated an adequate effect size (ES = 0.61). Change in scores for all scales in both disease groups were positive, indicating less disability on discharge than admission. Effect sizes on the MBI were similar to those of the FIM in both patient groups.

Hsueh, Lee and Hsieh (2001) examined the responsiveness of the MBI in 121 patients with stroke. The MBI was administered at 14, 30, 90 and 180 post-stroke. Standardized effect size scores were calculated for the intervals between 14-30 days, 30-90 days, 90-180 days and 14-180 days. The MBI demonstrated moderate to large effect sizes for all intervals, except for the 90-180 days post-stroke interval (ES = 0.56, 0.53, 0.11 and 1.27 respectively). The largest effect size was 14-180 days post-stroke, indicating that the MBI is most sensitive to detecting change in ADL function over longer periods of time.

Dromerick, Edwards, and Diringer (2003) examined responsiveness of the MBI and the FIM in a sample of 95 patients with stroke on admission to and discharge from a stroke rehabilitation service. The Modified Rankin Scale and the International Stroke Trial Measure were used to measure disability. The FIM was found to be more responsive to change from admission to discharge than the MBI, as calculated using the standardized response mean (SRM) (SRM= 2.18 vs. 1.72). The MBI detected change in 71/95 subjects but demonstrated ceiling effects with 27% of subjects scoring >95. The results of this study found the FIM to be the most sensitive of the four measures, detecting change in 91/ 95 patients, including change in 18 patients in whom the MBI detected no change.

Schepers, Ketelaar, Visser-Meily, Dekker and Lindeman (2006) investigated the responsiveness of the MBI, FIM, Frenchay Activities Indice (FAI), and Stroke Adapted Sickness Impact Profile 30 (SA-SIP30). The four measures were administered to 163 patients with stroke at admission to inpatient rehabilitation and at 6-months and 1-year post stroke. The MBI and the FIM total and motor scores were found to have a large effect sizes at 6-months post stroke (ES 0.98, 0.84 and 0.89 respectively) and a moderate effect size at 1-year post stroke (ES = 0.52, 0.47 and 0.51 respectively). The FIM cognitive score was found to have a moderate effect size at both 6-months and 1-year post stroke (ES = 0.47 at both time points). The SASIP30 and FAI demonstrated moderate effect sizes at 1-year post stoke (ES = 0.63 and 0.59 respectively). Results of this study indicate that the MBI and FIM (total and motor) are most apt to detect change in the subacute phase.
Note: The effect sizes for the SIP30 and FAI were not calculated at 6-months post stroke due to insufficient data. The FAI was only administered to patients who resided at home during the time of testing as the measure pertains to function relating daily housekeeping and activities typically performed outside of the rehabilitation or hospital environment.

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See the measure

For a copy of the original BI click here

Table of contents

DOC Screen

Evidence Reviewed as of before: 30-04-2019
Author(s)*: Alexandra Matteau
Editor(s): Annabel McDermott
Content consistency: Gabriel Plumier

Purpose

The DOC screen is a screening tool that can be used to identify individuals at high risk of depression, obstructive sleep apnea and cognitive impairment following a stroke.

In-Depth Review

Purpose of the measure

The DOC screen is a screening tool that identifies individuals at high risk of depression, obstructive sleep apnea and cognitive impairment following a stroke.

Available versions

The DOC screen was developed by Swartz et al. and was first published in 2013. The tool was developed by combining and modifying three existing validated brief screens, the 2-item Patient Health Questionnaire (PHQ-2), the STOP questionnaire and a 10-point version of the Montreal Cognitive Assessment (MoCA).

Features of the measure

Items:

The DOC screen comprises three screening tests:

DOC – Mood (PHQ-2)

This test comprises two items with the purpose of screening for depression. The test evaluates the degree to which an individual has experienced depressed mood and anhedonia over the past two weeks.

DOC – Apnea (STOP Questionnaire)

This test comprises four items with the purpose of screening for obstructive sleep apnea: snoring, tiredness during daytime, breathing interruption during sleep, and hypertension.

DOC – Cog (10-point version of the MoCA)

This test comprises three tasks with the purpose of screening for cognitive impairment: clock drawing, abstraction, and 5-word recall (memory).

Scoring:

Each subscale has different scoring and is interpreted independently.

DOC – Mood (total score 0-6)

The two items are scored from 0-3 whereby the respondent is asked to rate how often each symptom occurred over the last 2 weeks:

  • 0 = not at all
  • 1 = several days
  • 2 = more than half of the days
  • 3 = nearly every day.

DOC – Apnea (total score 0-4)

The four items are scored on a dichotomic scale (0 = no, 1 = yes) according to whether or not the respondent experiences each symptom.

DOC – Cog (total score 0-10)

  • Clock drawing task (0-3 points): 1 point each is given for (i) contour, (ii) numbers and (iii) the hands of the clock.
  • Abstraction task (0-2 points): 1 point is given for each item pair correctly answered.
  • Delayed recall task (0-5 points): 1 point is given for each word recalled without any cues.

The score for each task is summed to calculate the subscale score.

Each subscale is then summed to obtain a total score ranging between 0 and 20.

A raw score interpretation and a regression interpretation can be obtained at http://www.docscreen.ca/.

Time:

The DOC screen takes approximately 5 minutes to complete.

Subscales:

The DOC screen is comprised of three subscales: DOC Mood, DOC Apnea and DOC Cog.

Equipment:

A pencil and the test form are needed to complete the DOC screen.

Training:

No training requirements have been reported. The DOC screen can be administered by any individual who is able to correctly follow the instructions, but must be interpreted by a qualified health professional.

Alternative forms of the DOC Screen:

An alternative version is available and uses different words for the memory and abstraction tasks. This version must be used if the patient has previously been exposed to the MoCA or DOC screen to minimize any learning effects associated with repeated administration.

The E-DOC screen is an electronic version of the tool, which is available through the DOC screen website. The E-DOC screen has not been validated.

Client suitability

Can be used with:

  • Patients with stroke.
  • The DOC screen may also be suitable for use among patients with other neurological and vascular disorders such as multiple sclerosis, Alzheimer’s disease, mild cognitive impairment, Parkinson’s Disease and traumatic brain injury. However, no study has been conducted with this population.

Should not be used with:

While no contraindications have been reported, some considerations must be made when completing the test:

  • A translator, family member or caregiver can provide translation for patients who do not speak English fluently;
  • Provide visual aid (e.g. glasses) for patients with visual loss;
  • Speak loudly and clearly for patients with reduced hearing;
  • Motor tasks such as the clock drawing activity may be difficult for patients with motor impairments – use sound clinical judgement for this task;
  • Use alternative communication strategies for patients with aphasia.

In what languages is the measure available?

English

Summary

What does the tool measure? Depression, obstructive sleep apnea and cognitive impairment following stroke.
What types of clients can the tool be used for? Patients with stroke.
Is this a screening or assessment tool? Screening.
Time to administer 5 minutes
Versions
  • DOC screen
  • E-DOC screen
  • A second version is available to minimize learning effects associated with repeated administration.
Languages The DOC screen is only available in English.
Measurement Properties
Reliability Internal consistency:
No studies have examined internal consistency of the DOC screen.

Test-retest:
No studies have examined test-retest reliability of the DOC screen.

Intra-rater:
No studies have examined intra-rater reliability of the DOC screen.

Inter-rater:
No studies have examined inter-rater reliability of the DOC screen.

Validity Criterion:
Concurrent:
No studies have examined concurrent validity of the DOC screen.

Predictive:
No studies have examined predictive validity of the DOC screen.

Construct:
Convergent/Discriminant:
No studies have examined convergent validity of the DOC screen.

Known groups:
No studies have examined known groups validity. However, one study examined the sensitivity and specificity and reported that the DOC screen is a valid measure that can reliably identify patients at high-risk of depression, obstructive sleep apnea and cognitive impairment.

Floor/Ceiling Effects No studies have examined the floor or ceiling effects of the DOC screen.
Does the tool detect change in patients? Not reported.
Acceptability The DOC screen is a standardized screening tool suitable for use with stroke patients.
Feasibility The measure is brief, easy to score and requires no formal training. A study on 1503 patients showed that 89% of participants completed the screen in 5 minutes or less.
How to obtain the tool?

The DOC screen is free to use for clinical and educational purposes.

The administration manual and forms are available online from the following website: http://www.docscreen.ca/

Psychometric Properties

Overview

We conducted a literature search to identify all relevant publications on the psychometric properties of the DOC screen in individuals with stroke. We identified only one study, which was published in part by the developers of the measure. More studies are required before definitive conclusions can be drawn regarding the reliability and validity of the DOC screen.

Floor/Ceiling Effects

No studies have examined the floor or ceiling effects of the DOC screen.

Reliability

Internal consistency:
No studies have examined the internal consistency of the DOC screen.

Test-retest:
No studies have examined the test-retest reliability of the DOC screen.

Inter-rater:
No studies have examined the inter-rater reliability of the DOC screen.

Intra-rater:
No studies have examined the intra-rater reliability of the DOC screen.

Validity

Criterion:

Concurrent:
No studies have examined the concurrent validity of the DOC screen.

Predictive:
No studies have examined the predictive validity of the DOC screen.

Construct:

Convergent/Discriminant:
No studies have examined the convergent validity of the DOC screen.

Known groups:
No studies have examined the known groups validity of the DOC screen.

Responsiveness

No studies have examined the responsiveness of the DOC screen.

Sensitivity and Specificity:

Swartz et al. (2017) examined the sensitivity and specificity of the DOC screen for detecting depression, obstructive sleep apnea and cognitive impairment using receiver operating characteristic (ROC), area under the curve analyses (AUC) and the two-cut point approach. DOC-Mood was compared with the Structured Clinical Interview for DSM Disorders (SCID-D) and excellent sensitivity (92%) and specificity (99%) was identified for detecting depression (AUC=0.898). DOC-Apnea was compared with results on polysomnography (PSG) and excellent sensitivity (95%) and specificity (96%) for detecting obstructive sleep apnea was identified (AUC=0.660). DOC-Cog was compared to a 30-minute neuropsychological tests protocol proposed by Hachinski et al. (2006) and excellent sensitivity (100%) and specificity (95%) for detecting cognitive impairment was identified (AUC=0.776).

References

  • Hachinski, V., Iadecola, C., Petersen, R. C., Breteler, M. M., Nyenhuis, D. L., Black, S. E., … & Vinters, H. V. (2006). National Institute of Neurological Disorders and Stroke–Canadian stroke network vascular cognitive impairment harmonization standards. Stroke, 37 (9), 2220-2241.
  • Swartz, R. H., Cayley, M. L., Lanctôt, K. L., Murray, B. J., Cohen, A., Thorpe, K. E., … & Herrmann, N. (2017). The “DOC” screen: Feasible and valid screening for depression, Obstructive Sleep Apnea (OSA) and cognitive impairment in stroke prevention clinics. PloS one, 12 (4), e0174451.

See the measure

How to obtain the DOC Screen?

The form and manual of administration are available online from the following website: http://www.docscreen.ca/

The Doc screen is free to use for clinical and educational purposes and therefore no permissions are required.

Table of contents

Frenchay Activities Index (FAI)

Evidence Reviewed as of before: 19-08-2008
Author(s)*: Lisa Zeltzer, MSc OT
Editor(s): Nicol Korner-Bitensky, PhD OT; Elissa Sitcoff, BA BSc
Content consistency: Gabriel Plumier

Purpose

The Frenchay Activities Index (FAI) is a measure of instrumental activities of daily living (IADL) for use with patients recovering from stroke. The FAI assesses a broad range of activities associated with everyday life. The benefit of the FAI is that while activities of daily living scales tend to focus on issues related to self-care and mobility (Holbrook & Skilbeck, 1983), the FAI provides a broader measurement of actual activities patients have undertaken in the recent past (Wade, Legh-Smith, & Langton, 1985).

In-Depth Review

Purpose of the measure

The Frenchay Activities Index (FAI) is a measure of instrumental activities of daily living (IADL) for use with patients recovering from stroke. The FAI assesses a broad range of activities associated with everyday life. The benefit of the FAI is that while activities of daily living scales tend to focus on issues related to self-care and mobility (Holbrook & Skilbeck, 1983), the FAI provides a broader measurement of actual activities patients have undertaken in the recent past (Wade, Legh-Smith, & Langton, 1985).

Available versions

The FAI was published by Margaret Holbrook and Clive E. Skilbeck in 1983.

Features of the measure

Items:

The FAI contains 15 items or activities that can be separated into 3 subscales; Domestic chores, Leisure/work and Outdoor activities.

The items of the FAI are as follows:

  1. Preparing main meals
    Must play a substantial part in organization, preparation and cooking.
  2. Washing up
    Must do all or share equally, e.g. washing or wiping and putting away.
  3. Washing clothes
    Organization of washing and drying clothes. Sharing task equally, e.g. loading, unloading, hanging, folding.
  4. Light housework
    Dusting, ironing, tidying small objects. Anything heavier is included in item 5.
  5. Heavy housework
    Changing beds, cleaning floors, windows, vacuuming, moving chairs, etc.
  6. Local shopping
    Substantial role in organizing and buying groceries. Can include collection of pension or going to the Post Office.
  7. Social outings
    Going out to clubs, cinema, theatre, drinking, dinner with friends, etc. May be transported there, provided patient takes an active part once arrived. Includes social activities at home, initiated by the patient.
  8. Walking outdoors over 15 minutes
    Sustained walking for at least 15 minutes (allowed short stops for breath).
  9. Pursuing active interest in hobby
    Must require ‘active’ participation, e.g. caring for houseplants, knitting, reading specialist magazines or window-shopping.
  10. Driving a car
    Must drive a car, or get to a bus/coach and travel on it independently.
  11. Outings/car rides
    Train, bus, or car rides to some place for pleasure, not for a routine social outing. Must involve patient organization and decision-making. Holidays within the last 6 months are divided into days/month (e.g. a 7-day holiday = 1 or 2 days/month).
  12. Gardening
    Light = occasional weeding or sweeping; Moderate = regular weeding, raking, pruning; Heavy = all necessary work including heavy digging.
  13. Household and/or car maintenance
    Light = repairing small items, replacing lightbulb or plug; Moderate = spring cleaning, hanging a picture, routine car maintenance; Heavy = painting/decorating, most necessary household/car maintenance.
  14. Reading books
    Full-length books, not magazines or newspapers. Can be talking books.
  15. Gainful work
    Paid work, not voluntary work. The time worked should be averaged out over six months (e.g., 1 month working for 18 hours/week over the 6-month period would be scored as ‘up to 10 hours/week’).

Time:

The FAI takes approximately 5 minutes to complete when administered in an interview format (with or without the patient’s family) (Segal & Schall, 1994).

Scoring:

The frequency with which each item or activity is undertaken over the past 3 or 6 months (depending on the nature of the activity) is assigned a score of 1 – 4 where a score of 1 = lowest level of activity. The scale provides a summed score from 15 – 60.

A modified 0-3 scoring system introduced by Wade et al. (1985) yields a score of 0 – 3 for each item, and a summed score from 0 – 45.

Note: In patients with stroke, the FAI should be used to assess pre-morbid IADL at 3 and 6 months before stroke, and subsequently to record changes in IADL following stroke, at specific intervals (Holbrook & Skilbeck, 1983). Studies typically examine change in post-stroke IADL by examining patients at 1 year after stroke, and looking retrospectively at the past 3 and 6 months.

Subscales:

There are 3 subscales to the FAI:

  • Domestic (items 1-5)
  • Leisure/work (items 7, 9, 11, 13, 15)
  • Outdoors (items 6, 8, 10, 12, 14)

Equipment:

Only the questionnaire and a pencil are needed to complete the FAI.

Training:

No training is required to complete the FAI. The FAI is most often interview-administered.

The FAI can be used as a mailed questionnaire. Carter, Mant, Mant, Wade, and Winner (1997) reported an excellent correlation between mailed questionnaire FAI scores and face-to-face interview scores (r = 0.94).

The FAI can also be used with a proxy respondent. Proxy agreement was excellent for the FAI (intraclass correlation coefficient (ICC) = 0.85) (Segal & Schall, 1994). Holbrook and Skilbeck (1983) found that information obtained by relatives were interchangeable with information acquired from the patient. Segal and Schall (1994) reported proxy agreement for the three subscales as ranging from adequate (ICC = 0.59 for Leisure/work) to excellent (ICC = 0.77 for Domestic and Outdoors).

Alternative Forms of the FAI

  • FAI-18 (Miller, Deathe, & Harris, 2004).
    Three items (sport/recreation and visiting in the last 3 months, and banking in the last 6 months) were added to the FAI and the reliability was examined in patients with lower limb amputation. The total score of the FAI-18 ranges from 0 to 54. Support for the concurrent validity (r = -0.46), the Prosthetic Evaluation Questionnaire-Mobility Scale (r = 0.40) and the Activities-specific Balance Confidence Scale (r = 0.52). The FAI-18 was not found to offer any advantage over the original FAI and therefore use of the original FAI is recommended to ensure results are comparable between populations and studies. Further, the FAI-18 has not been examined in patients with stroke.
  • Modified FAI (Tooth, McKenna, Smith, & O’Rourke, 2003).
    A 13-item modified version has been developed based on the recommendations by Schuling, de Haan, Limburg, and Groenier (1993) to omit the items ‘reading books’ and ‘gainful work’. At 6 months post-stroke, the internal consistency of the 13 FAI items was excellent when scored by patients (alpha = 0.85) and when scored by proxies (alpha = 0.83). However, the internal consistency of each subscale examined separately varied widely.

Client suitability

Can be used with

  • Patients with stroke.
  • Can also be used with patients with cognitive impairment, using a proxy respondent. The focus of the FAI is on frequency of activity rather than quality of activity. This may reduce elements of subjectivity, which typically undermine the reliability of proxy assessment (Segal & Schall, 1994).

Should not be used with

  • When examining FAI scores, male and female scores should be considered separately as there is evidence of a gender bias in FAI scores (Holbrook & Skilbeck, 1983). Sveen, Bautz-Holter, Sodring, Wyller, and Laake (1999) reported that men had significantly higher scores in the Outdoor activities subscale, and there was a trend towards women having higher scores in the Domestic activity subscale.
  • Due to individual variability, the FAI should not be administered by interview and by mailed questionnaire, sequentially (Carter et al., 1997).
  • Use caution when examining proxy ratings at the item level, because there is less agreement than what has been observed with the total score (Wyller, Sveen, & Bautz-Holter, 1996; Tooth et al., 2003).
  • Be aware of the biases involved with proxy use. Tooth et al. (2003) reported that patients tend to score themselves as performing activities more frequently than proxy respondents especially in meal preparation, heavy housework, social outings, driving and home maintenance. In addition, male proxy respondents and respondents who are relatives (rather than spouses) tend to give higher ratings, particularly in the area of domestic activities.

In what languages is the measure available?

  • English
  • Dutch – translated (Schuling, de Haan, Limburg, & Groenier, 1993)
  • Chinese – translated and validated (Hsueh & Hsieh, 1997)

Summary

What does the tool measure? Instrumental Activities of Daily Living
What types of clients can the tool be used for? Patients with stroke
Is this a screening or assessment tool? Assessment
Time to administer Interview: 5 minutes (with or without the patient’s family)
Versions FAI-18, Modified FAI
Other Languages Chinese (translated and validated), Dutch (translated)
Measurement Properties
Reliability Internal consistency:
Out of three studies examining internal consistency, three reported excellent internal consistency.

Test-retest:
Out of four studies examining test-retest, three reported excellent test-retest, and one reported a range from poor to excellent depending on item examined.

Inter-rater:
Out of two studies examining inter-rater reliability, two studies reported excellent inter-rater reliability as measured by intraclass correlation coefficients. Using Cohen’s kappa, one study reported adequate to excellent reliability and one study reported poor to excellent reliability.

Validity Content:
Three studies examined the content validity of the FAI suggesting the presence of a single underlying construct in that each item contributes to each of the three identified factors (Domestic; Leisure/work; Outdoors)

Criterion:
Excellent correlation between postal and interview FAI scores, however individual differences on scores ranged widely between mailed and postal responses taken 10 days later.

Construct:
Excellent correlations with Rankin Scale ; SF-36 (Physical Functioning subscale). Adequate to excellent correlations with the Sickness Impact Profile ; Barthel Index ; Functional Independence Measure (Motor subscale); Euroqol. Adequate correlations with Stroke Adapted Sickness Impact Profile ; SF-36 (Social Functioning and Vitality subscales); two-minute walk test, Timed Up and Go test ; Prosthetic Evaluation Questionnaire-Mobility; Activities-specific Balance Confidence Scale.

Known groups:
The FAI has been found to distinguish stroke severity in male patients only and can discriminate between patients in a pre-stroke versus a reference group, and patients’ pre-stroke and post-stroke levels of activity.

Does the tool detect change in patients? One study reported an “obvious” floor effect for individuals examined at 6 months post-stroke.

Out of two studies examined, one reported that the FAI had a moderate ability to detect change (in patients 6-12 months post-stroke) and one reported that the FAI changed in the expected direction from pre-stroke to 6 months post-stroke, to 1 year post-stroke.

Acceptability The FAI is short, simple, and encourages participation of significant others or family members. It is suitable for use with proxy respondents.
Feasibility The FAI is simple to administer and requires no training or special equipment. It has been used for longitudinal assessment.
How to obtain the tool? A copy of the original FAI provided in Holbrook, M., Skilbeck, C. E. (1983). An activities index for use with stroke patients. Age and Ageing, 12(2), 166-170.

Psychometric Properties

Overview

For the purposes of this review, we conducted a literature search to identify all relevant publications on the psychometric properties of the FAI. In general, the FAI has good overall reliability, however it has considerable variability in the strength of agreement at the level of individual scale item scores (reported both for test-retest and inter-rater reliability). Further, there is little evidence regarding the responsiveness of the FAI.

Floor/Ceiling Effects

Schuling et al. (1993) examined the psychometric properties of the FAI in a group of patients with stroke and a control group of individuals from the general population aged 65 or older. No ceiling effects were reported in this study.

Similarly, Wade et al. (1985) examined the psychometric properties of the FAI using data from 976 patients with acute stroke. No ceiling effects were reported.

Pederson et al. (1997) examined the FAI in 437 patients with stroke and reported an “obvious” floor effect at 6 months post-stroke.

Walters, Morrell and Dixon (1999) examined the psychometric properties of four generic instruments in 233 patients with venous leg ulcers. The FAI demonstrated an adequate floor effect of 2.1%. No ceiling effect was observed.

Reliability

Internal consistency:

Schuling et al. (1993) examined the internal consistency of the FAI retrospectively in a group of patients with stroke and a control group of individuals from the general population aged 65 or older. They looked at the internal consistency of the FAI pre-stroke, 6 months post-stroke and in control patients. An excellent internal consistency was reported for the total score of the FAI in the control group (alpha = 0.83) and in patients post-stroke (alpha = 0.87). An adequate alpha coefficient was reported for patients pre-stroke (alpha = 0.78). When subscales were examined individually, the Domestic subscale had excellent alpha coefficients (alpha = 0.82 for control and pre-stroke; 0.88 for post-stroke). The Leisure/work subscale had poor internal consistency in all groups (control, alpha = 0.63; pre-stroke, alpha = 0.58; post-stroke, alpha = 0.61). The Outdoors subscale also had poor internal consistency in all groups (control, alpha = 0.67; pre-stroke, alpha = 0.55; post-stroke, alpha = 0.66). However, when item 14 (reading books) was deleted, alpha coefficients were adequate for control and post-stroke groups (alpha = 0.72, alpha = 0.73, respectively) and remained poor in the pre-stroke group (alpha = 0.66).

Tooth et al. (2003) examined the agreement between patients with stroke and their proxies using a modified version of the FAI (13 items). At 6 months post-stroke, the internal consistency of the 13 FAI items was excellent when scored by patients (alpha = 0.85) and when scored by proxies (alpha = 0.83). The internal consistency of each subscale examined separately varied widely. Coefficient alphas for the Domestic, Leisure, and Outdoor subscales completed by patients ranged from poor to excellent (0.83, 0.38, 0.66, respectively), as did completion by proxies (0.83, 0.59, 0.57, respectively).

Miller et al. (2004) compared the reliability of the FAI to a modified version, the FAI-18. The internal consistency of the FAI was excellent (alpha = 0.81).

Test-retest:

Wade et al. (1985) examined the test-retest reliability of the FAI and reported that the overall agreement of individual items was variable. Heavy housework, local shopping, walking outside and social outings failed to reach statistical significance, while other items demonstrated excellent agreement (r = 0.80).

Green, Forster, and Young (2001) examined the test-retest reliability of the Barthel Index (Mahoney & Barthel, 1965), the Rivermead Mobility Index (Nouri & Lincoln, 1987), the Nottingham Extended Activities of Daily Living Scale (Whiting & Lincoln, 1980), and the FAI in 22 patients > 1 year post-stroke, tested twice at an interval of 1 week. Kappa coefficients for the FAI ranged from poor (kappa = 0.25 for heavy housework) to excellent (kappa = 1.00 for preparing main meals). The results of this study indicate that basic measures of activities of daily living (as measured by the Barthel Index and Rivermead Mobility Index) may be more reliable than the measures used to assess IADL.

Turnbull, Kersten, Habib, McLellan, Mullee, and George (2000) assessed the reliability of the FAI to establish age and sex norms in people age 16 years and over. A postal questionnaire survey was sent to 1,280 people. Then 57 respondents completed a re-test questionnaire. Test-retest reliability of the postal version of the FAI was excellent, with a correlation of r = 0.96.

Miller et al. (2004) examined the reliability of the FAI in 84 individuals with lower limb amputation. Individuals completed the FAI twice, within two weeks. The ICC for the FAI was excellent (ICC = 0.79), demonstrating the test-retest reliability of the FAI.

Inter-rater:

Piercy, Carter, Mant, and Wade (2000) examined the inter-rater reliability of the FAI in 35 patients with stroke and 24 individuals who were the main caregivers for patients with stroke. Two raters evaluated each person, 15 days apart on average. Kappa statistics showed an excellent level of agreement for 3/15 items (kappas ranging from 0.77-0.80). An adequate level of agreement was found for 10/15 items (kappas ranging from 0.42-0.73). The other 2 items showed poor agreement (social outings, 0.27; pursuing active interest in hobby, 0.35). Three items showed significant differences between the two raters (light housework, outing/car rides, household and/or car maintenance). Spearman’s correlation for FAI totals of rater B verses rater A was excellent (r = 0.93). The results of this study confirm the reliability of the FAI when administered by interview.

Post and de Witte (2003) examined the inter-rater reliability of the Dutch version of the FAI in 45 patients with stroke. The FAI was administered twice, with 3-5 days in between evaluations. The total inter-rater reliability of the FAI was excellent (ICC = 0.90). At item level, kappa coefficients ranged from adequate to excellent (kappa = 0.41-0.90).

Validity

Content:

Wade et al. (1985) examined data from 976 patients with acute stroke. A factor analysis was conducted to demonstrate levels of communality among the FAI’s items. Correlations ranged from 0.44-0.77, suggesting the presence of a single underlying construct in that each item contributes to each of the three identified factors (Domestic; Leisure/work; Outdoors) to some extent.

Pedersen, Jorgensen, Nakayama, Raaschou, and Olsen (1997) examined whether the FAI was a good supplementary assessment to the Barthel Index (Mahoney & Barthel, 1965) for measuring higher order ADL functions in 437 patients with stroke. The FAI was found to be a heterogeneous scale comprised of 3 factors, two of which may represent increased item difficulties, and the third related to activities away from the home. Items from the Barthel Index and the FAI, when analyzed together, appeared on different, orthogonal factors, suggesting that the FAI supplements the Barthel Index with minimal content overlap.

Sveen et al. (1999) examined data from 65 patients with stroke to observe how motor and cognitive impairments relate to physical activities of daily living. In this study, the 3-factor structure of the FAI was confirmed. These three subscales include Domestic chores, Outdoor activities and Hobbies.

Criterion:

Concurrent:
Carter et al. (1997) examined the agreement between postal and interview-administered versions of the FAI, and assessed the criterion validity of the postal version, using the interviewer method as the gold standard. An excellent Spearman’s correlation of r = 0.94 was found between mailed questionnaire FAI scores and face-to-face interview FAI scores. Individual differences on scores ranged widely between FAI responses by post and responses by interview 10 days later. At the level of individual items, kappas ranged from poor (kappa = 0.35 for travel outings/car rides) to excellent (kappa = 1.00 for gainful work). The postal version was found to be a satisfactory alternative to interview administration, however, due to poor agreement in scores for individual patients, the two approaches should not be used sequentially to monitor individual patient.

Cup, Scholte op Reimer, Thijssen, and van Kuyk-Minis (2003) administered a number of different standardized measures to 26 patients with stroke. The FAI had excellent correlations with the Barthel Index (Mahoney & Barthel, 1965) (r = 0.79), the Euroqol (r = 0.65) (EuroQol Group, 1990), and the Rankin Scale (r = -0.80) (de Haan, Limburg, Bossuyt, van der Meulen, & Aaronson, 1995). The FAI an adequate correlation with the Stroke Adapted Sickness Impact Profile-30 (van Straten, de Haan, Limburg, Schuling, Bossuyt, & van den Bos, 1997) (r = -0.43).
Note: Some correlations are negative because a high score on the FAI indicates a high level of functioning, where as a high score on the Stroke Adapted Sickness Impact Profile-30 and the Rankin Scale indicates less desirable health outcomes.

Segal and Schall (1994) examined the proxy agreement between 38 patients with stroke and their caregivers. Using Spearman’s rho, the FAI and the Functional Independence Measure (Keith et al., 1987) were found to have an excellent correlation (r = 0.80).

Hsueh, Lee, and Hsieh (2001) examined the psychometric properties of the Barthel Index (Mahoney & Barthel, 1965) in 121 patients with stroke. The FAI was compared to the Barthel Index at 180 days after stroke and was found to have an adequate correlation with the Barthel Index scores obtained at 14, 30, and 90 days after stroke (Pearson’s r = 0.59).

Walters et al. (1999) examined the psychometric properties of four generic instruments: Short-Form Health Survey (SF-36) (Ware & Sherbourne, 1992); EuroQol (EuroQol Group, 1990); McGill Short Form Pain Questionnaire (Melzack, 1975) and the FAI in 233 patients with venous leg ulcers. Correlations were calculated using Pearson Product Moment Correlations. The FAI had an excellent correlation with the SF-36 subscale of Physical Functioning (r = 0.72). Poor correlations between FAI and the SF-36 subscales of Role Limitations-Physical (r = 0.25), Role Limitations-Emotional (r = 0.11), Pain (r = 0.28), General Health Perceptions (r = 0.30), and Mental Health (r = 0.26) were observed. Adequate correlations were found between the FAI and the SF-36 subscales of Social Functioning (r = 0.35), and Vitality (r = 0.45). The FAI had a moderate correlation with the EuroQol Derived Single Index (r = 0.54), and a poor correlation with the McGill Pain Questionnaire Sensory (r = -0.12) and Affective (r = -0.13) subscales. Note: Some correlations are negative because a high score on the FAI indicates a high level of functioning, where as a high score on other measures indicates less desirable health outcomes.

Miller et al. (2004) examined the concurrent validity of the FAI in 84 individuals with lower limb amputation. As predicted, the FAI correlated adequately with the Two-minute walk test (r = 0.53), the Timed Up and Go test (Podsiadlo & Richardson, 1991) (r = -0.49), the Prosthetic Evaluation Questionnaire-Mobility Scale (Legro, Reiber, Smith, del Aguila, Larsen, & Boone, 1998) (r = 0.39), and the Activities-specific Balance Confidence Scale (Powell & Myers, 1995) (r = 0.51).
Note: Some correlations are negative because a high score on the FAI indicates a high level of functioning, whereas a high score on the Timed Up and Go test indicates less desirable health outcomes.

Construct:

Convergent/Discriminant:
Schuling et al. (1993) examined the construct validity of the FAI in patients with stroke, and in a group of unselected participants aged 65 or older. Functional status of the patients with stroke was measured at 26 weeks. Correlations between the FAI and the Sickness Impact Profile (Bergner, Bobbitt, Carter, & Gilson, 1981) subscales of Home management, Body care and movement, Mobility and Ambulation ranged from adequate to excellent (r = -0.56 to -0.73). The FAI also had an excellent correlation with the disability scores of the Barthel Index (Wade & Collin, 1988) (r = 0.66). These results provide evidence for the convergent validity of the FAI. Further, the discriminant validity of the FAI is supported by the poor correlation found between FAI scores and Emotional Behavior and Alertness Behavior scales of the Sickness Impact Profile (r = -0.15 and -0.14).
Note: Some correlations are negative because a high score on the FAI indicates a high level of functioning, where as a high score on the Sickness Impact Profile indicates less desirable health outcomes.

Sveen et al. (1999) examined data from 65 patients with stroke and found that Domestic chores and Outdoor activities (factors found in this study to make up the FAI) correlated adequately with Barthel Index (Mahoney & Barthel, 1965) scores (r = 0.58 and r = 0.50). Domestic chores was the factor most strongly related to arm motor function of the Barthel Index, and Outdoor activities was most strongly related to visuospatial ability. Hobbies, the third factor found in this study, did not correlate with Barthel Index scores (r = 0.11).

Tooth et al. (2003) examined the construct validity of the FAI in patients with stroke and their proxies using a modified version of the index (13 items). The total patient FAI score was found to correlate significantly with the Motor subscale of the Functional Independence Measure (Keith, Granger, Hamilton, & Sherwin, 1987) (r = 0.63) but not with the Cognitive subscale of the Functional Independence Measure (r = 0.09).

Known groups:
Holbrook and Skilbeck (1983) divided patients by stroke severity into ‘mild’ and ‘severe’ based on Rankin grade at the time of stroke. They reported that the FAI distinguished severity of stroke (by Rankin groupings) in male patients, who showed significantly poorer Domestic chores scores and Outdoor activities scores at follow up. However, stroke severity did not influence one-year follow-up for females.

Schuling et al. (1993) reported that the FAI was able to discriminate between patients in the pre-stroke group and patients in the reference group. The FAI was also discriminative of patients’ pre-stroke and post-stroke levels of activity.

Responsiveness

Schepers, Ketelaar, Visser-Meily, Dekker, and Lindeman (2006) compared the responsiveness of frequently used functional health status measures in stroke. The FAI and the Stroke Adapted Sickness Impact Profile detected the most changes and had moderate effect sizes for patients in the chronic phase (between 6 and 12 months post-stroke) of stroke rehabilitation.

Wade et al. (1985) reported that FAI scores changed in the expected direction from pre-stroke to 6 months post-stroke to 1 year post-stroke.

References

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  • Hsueh, I.-P., Hsieh, C.-L. (1997). A revalidation of the Frenchay Activities Index in stroke: A study in Taipei area. Formorsan Med J, 6, 123-130 [in Chinese].
  • Keith, R. A., Granger, C. V., Hamilton, B. B., Sherwin, F. S. (1987). The functional independence measure: A new tool for rehabilitation. Adv Clin Rehabil, 1, 6-18.
  • Legro, M. W., Reiber, G. D., Smith, D. G., del Aguila, M., Larsen, J., Boone, D. (1998). Prosthesis Evaluation Questionnaire for persons with lower limb amputations: assessing prothesis-related quality of life. Arch Phys Med Rehabil, 79, 931-938.
  • Mahoney, F. I., Barthel, D. W. (1965). Functional evaluation: The Barthel Index. Md State Med J, 14, 61-65.
  • Melzack, R. (1975). The McGill Pain Questionnaire: Major Properties and Scoring Methods. Pain, 1, 277-289.
  • Miller, W. C., Deathe, A. B., Harris, J. (2004). Measurement properties of the Frenchay Activities Index among individuals with a lower limb amputation. Clinical Rehabilitation, 18(4), 414-422.
  • Nouri, F. M., Lincoln, N. B. (1987). An extended activities of daily living scale for stroke patients. Clin Rehab, 1, 301-305.
  • Pedersen, P. M., Jorgensen, H. S., Nakayama, H., Raaschou, H. O., Olsen, T. S. (1997). Comprehensive assessment of activities of daily living in stroke. The Copenhagen Stroke Study. Arch Phys Med Rehabil, 78, 161-165.
  • Piercy, M., Carter, J., Mant, J., Wade, D. T. (2000). Inter-rater reliability of the Frenchay Activities Index in patients with stroke and their carers. Clinical Rehabilitation, 14, 433-440.
  • Podsiadlo, E., Richardson, S. (1991). The Timed ‘Up & Go’: a test of basic functional mobility for frail elderly persons. J Am Geriatr Soc, 39, 142-148.
  • Post, M. W. M., de Witte, L. P. (2003). Good inter-rater reliability of the Frenchay Activities Index in stroke patients. Clinical Rehabilitation, 17(5), 548-552.
  • Powell, L., Myers, A. (1995). The Activities-specific Balance Confidence (ABC) scale. J Gerontol, 50, M28-34.
  • Schepers, V. P. M., Ketelaar, M., Visser-Meily, J. M. A., Dekker, J., Lindeman, E. (2006). Responsiveness of functional health status measures frequently used in stroke research. Disability & Rehabilitation, 28(17), 1035-1040.
  • Schuling, J., de Haan, R., Limburg, M., Groenier, K. H. (1993). The Frenchay Activities Index. Assessment of functional status in stroke patients. Stroke, 24, 1173-1177.
  • Segal, M. E., Schall, R. R. (1994). Determining functional/health status and its relation to disability in stroke survivors. Stroke, 25, 2391-2397.
  • Sveen, U., Bautz-Holter, E., Sodring, K. M., Wyller, T. B., Laake, K. (1999). Association between impairments, self-care ability and social activities 1 year after stroke. Disability & Rehabilitation, 21(8), 372-377.
  • The EuroQol Group. (1990). EuroQol – a facility for the measurement of health-related quality of life. Health Policy, 16, 199-207.
  • Tooth, L. R., McKenna, K. T., Smith, M., O’Rourke, P. (2003). Further evidence for the agreement between patients with stroke and their proxies on the Frenchay Activities Index. Clinical Rehabilitation, 17, 656-665.
  • Turnbull, J. C., Kersten, P., Habib, M., McLellan, L., Mullee, M. A., George, S. (2000). Validation of the Frenchay Activities Index in a general population aged 16 years and older. Arch Phys Med Rehabil, 81(8), 1034-1038.
  • van Straten, A., de Haan, R. J., Limburg, M., Schuling, J., Bossuyt, P. M., van den Bos, G. A. M. (1997). A Stroke-Adapted 30-Item Version of the Sickness Impact Profile to Assess Quality of Life (SA-SIP30). Stroke, 28, 2155-2161.
  • Wade, D. T., Legh-Smith, J., Langton, H. R. (1985). Social activities after stroke: Measurement and natural history using the Frenchay Activities Index. Int Rehabil Med, 7(4), 176-181.
  • Ware, J. E. Jr., Sherbourne, C. D. (1992). The MOS 36-item short-form health survey (SF-36). I. Conceptual framework and item selection. Med Care, 30, 473-483.
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  • Wade, D. T., Collin, C. (1988). The Barthel ADL Index: a standard measure of physical disability. Int Disability Studies, 10, 64-67.
  • Walters, S. J., Morrell, J., Dixon, S. (1999). Measuring health-related quality of life in patients with venous leg ulcers. Quality of Life Research, 8, 327-336.
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See the measure

How to obtain the FAI:

For a copy of the FAI with the scoring system by Wade et al. (1985), please click here.

A copy of the measure with the original scoring system is also provided in Holbrook, M., Skilbeck, C. E. (1983). An activities index for use with stroke patients. Age and Ageing, 12(2), 166-170.

Table of contents

Functional Independence Measure (FIM)

Evidence Reviewed as of before: 15-10-2011
Author(s)*: Lisa Zeltzer, MSc OT;
Editor(s): Nicol Korner-Bitensky, PhD OT; Elissa Sitcoff, BA BSc

Purpose

The Functional Independence Measure (FIM) was developed to address the issues of sensitivity and comprehensiveness that were criticized as being problematic with the Barthel Index (another measure of functional independence). The FIM was also developed to offer a uniform system of measurement for disability based on the International Classification of Impairment, Disabilities and Handicaps for use in the medical system in the United States (McDowell & Newell, 1996). The level of a patient’s disability indicates the burden of caring for them and items are scored on the basis of how much assistance is required for the individual to carry out activities of daily living.

In-Depth Review

Purpose of the measure

The Functional Independence Measure (FIM) was developed to address the issues of sensitivity and comprehensiveness that were criticized as being problematic with the Barthel Index (another measure of functional independence). The FIM was also developed to offer a uniform system of measurement for disability based on the International Classification of Impairment, Disabilities and Handicaps for use in the medical system in the United States (McDowell & Newell, 1996). The level of a patient’s disability indicates the burden of caring for them and items are scored on the basis of how much assistance is required for the individual to carry out activities of daily living.

The FIM assesses six areas of function (Self-care, Sphincter control, Transfers, Locomotion, Communication and Social cognition), which fall under two Domains (Motor and Cognitive). It has been tested for use in patients with stroke, traumatic brain injury, spinal cord injury, multiple sclerosis, and elderly individuals undergoing inpatient rehabilitation and has been used with children as young as 7 years old.

Available versions

The FIM was developed between 1984 and 1987 by a national task force sponsored by the American Academy of Physical Medicine and Rehabilitation and the American Congress of Rehabilitation Medicine and was published by Keith, Granger, Hamilton, and Sherwin in 1987.

Features of the measure

Items:

The FIM consists of 18 items assessing 6 areas of function. The items fall into two domains: Motor (13 items) and Cognitive (5 items). The motor items are based on the items of the Barthel Index. These domains are referred to as the Motor-FIM and the Cognitive-FIM.

The items of the FIM are listed as follows:

Motor Domain:

1. Self-care (6 items)

– Eating
– Grooming
– Bathing
– Dressing – Upper body
– Dressing – Lower body
– Toileting

2. Sphincter control (2 items)

– Bladder management
– Bowel management

3. Transfers (3 items)

– Bed/Chair/Wheelchair
– Toilet
– Tub/Shower

4. Locomotion (2 items)

– Walk/Wheelchair
– Stairs

Cognitive Domain:

5. Communication (2 items)

– Comprehension
– Expression

6. Social cognition (3 items)

– Social interaction
– Problem solving
– Memory

For the Motor-FIM, the Eating, Grooming, and Bowel management items are known to be the easiest items for patients with stroke to accomplish, whereas Tub/Shower transfers and Locomotion (Walk/Wheelchair, Stairs) are the most challenging items (Granger, Cotter, Hamilton, & Fiedler, 1993; Grimby, Gudjonsson, Rodhe, Sunnerhagen, Sundh, & Ostensson, 1996). For the Cognitive-FIM, performance of the Expression item has been found to be the easiest for patients to accomplish, and Problem solving has been found to be the most challenging (Granger et al., 1993).

Time:

The FIM is reported to take between 30-45 minutes to administer and score, with 7 minutes to gather demographic information.

Scoring:

Each item on the FIM is scored on a 7-point Likert scale, and the score indicates the amount of assistance required to perform each item (1 = total assistance in all areas, 7 = total independence in all areas). The ratings are based on performance rather than capacity and can be acquired by observation, patient interview, telephone interview or medical records. The developers of the FIM recommend that the scoring be derived by consensus with a multi-disciplinary team.

A final summed score is created and ranges from 18 – 126, where 18 represents complete dependence/total assistance and 126 represents complete independence. The single summed raw score may be misleading as it gives the appearance of a continuous scale. However, intervals between scores are not equal in terms of level of difficulty and cannot provide more than ordinal level information (Linacre et al., 1994). Kidd et al. (1995) suggested using the summed scores as though on an interval scale while the individual items remain ordinal. Granger, Deutsch, and Linn (1998) have applied a Rasch rating scale in order to transform the FIM’s ordinal ratings to an equal-interval rating scale so that it can be used for linear regression models.

Subscale scores for the Motor and Cognitive domains can also be calculated (Linacre, Heinemann, Wright, Granger, & Hamilton, 1994).

Equipment:

Any items that the patient uses to carry out their activities of daily living.

Subscales:

There are two subscales for the FIM: the Motor-FIM and the Cognitive-FIM.

Training:

The FIM must be administered by a trained and certified evaluator.

Grey and Kennedy (1993) found that the FIM could be completed as a self-report questionnaire in patients with spinal cord injury. Segal and Schall (1994) found that the FIM can be used reliably by in-person proxy for patients with stroke. Segal, Gillard, and Schall (1996) further established that the FIM can be used reliably by proxy over the telephone in patients with stroke (total FIM, intraclass correlation coefficient (ICC) = 0.91, Motor-FIM, ICC = 0.94; Cognitive-FIM, ICC = 0.52), and closely resembles results obtained for the in-person administration.

Alternative Forms of the Functional Independence Measure

  • The Functional Independence Measure for Children (WeeFIM). This measure was developed to track disability in children who are between the ages of 6 months and 7 years. The WeeFIM can be administered to children over the age of 7 if their functional abilities are below those expected of children aged 7 who do not have disabilities. It measures the impact of developmental strengths and difficulties on independence at home, in school, and in the community (Msall et al., 1994). The scale has 18 items measuring functional performance in 3 domains: Self-care, Mobility, and Cognition (Uniform Data System for Medical Rehabilitation, http://www.udsmr.org/).
  • Modified 5-level FIM. Gosman-Hedström and Blomstrand (2004) examined whether a 5-level FIM would be more useful than the standard 7-level version in large population studies. They used a sample of elderly stroke survivors and found that a 5-level FIM would most likely increase the reliability of the FIM without losing sensitivity.

Client suitability

Can be used with:

Patients with stroke of all ages, and can be used with patients with special conditions (e.g. aphasia or neglect).

Should not be used in:

No restrictions have been reported.

In what languages is the measure available?

The FIM has been translated in the following languages:

  • German
  • Italian
  • Spanish
  • Swedish
  • Finnish
  • Portuguese
  • Afrikaans
  • Turkish
  • French
  • Persian (Farsi)

Summary

What does the tool measure? Activities of Daily Living
What types of clients can the tool be used for? Patients with stroke, traumatic brain injury, spinal cord injury, multiple sclerosis, and elderly individuals undergoing inpatient rehabilitation. Can be used with children as young as 7 years old.
Is this a screening or assessment tool? Assessment
Time to administer The FIM is reported to take between 30-45 minutes to administer and score, with 7 minutes to gather demographic information.
Versions WEE-FIM; Modified 5-level FIM
Other Languages German; Italian; Spanish; Swedish; Finnish; Portuguese; Afrikaans; Turkish; French; Persian (Farsi).
Measurement Properties
Reliability Internal consistency:
Out of four studies examining internal consistency, all four reported excellent internal consistency.

Test-retest:
Out of five studies examining test-retest reliability, all five reported excellent test-retest.

Inter-rater:
Out of 10 studies examining inter-rater reliability, eight studies reported excellent; one reported adequate to excellent (except Social Interaction item which was poor); one reported overall poor kappa values, but excellent intraclass correlation coefficient (ICC).

Validity Content:
The FIM was created based on the results of a literature review of published and unpublished measures and expert panels and was then piloted in 11 centers. The Delphi method was applied, using rehabilitation expert opinion to establish the inclusiveness and appropriateness of the items.

Criterion:
Excellent correlations with the Barthel Index; Modified Rankin Scale; Disability Rating Scale. FIM scores found to predict amount of home care required; admission scores predict FIM discharge scores; placement after discharge; functional gain; length of stay; depression, ability to return to work following stroke or traumatic brain injury.

Concurrent:
The Motor-FIM was found to demonstrate an excellent correlation with the Modified Rankin Scale (MRS) and the Disability Rating Scale (DRS); and adequate to excellent correlation with the Barthel Index. The Cognition-FIM was found to have an excellent correlation with the DRS; an adequate correlation with the Montebello Rehabilitation Factor Score (MRFS) (efficacy); and a poor correlation with the MRFS (efficiency).

Construct:
FIM scores discriminated between groups based on spinal cord injury and stroke severity, and the presence of comorbid illness both at admission and discharge. It has also been found to distinguish between patients with or without neglect and with or without aphasia at both admission and discharge.

Convergent/Discriminant:
The total FIM was found to demonstrate an excellent correlation with the Office of Population Censuses and Surveys Disability Scales disability scores; an adequate correlation with the London Handicap Scale and the Wechsler Adult Intelligence Test-verbal IQ test; and a poor correlation with the SF-36 Physical and Mental components, and the General Health Questionnaire. The Motor-FIM was found to demonstrate an excellent correlation with the Office of Population Censuses and Surveys Disability Scales disability scores; an adequate correlation with the London Handicap Scale; and a poor correlation with the Wechsler Adult Intelligence Test-verbal IQ test, SF-36 Physical and Mental components, and the General Health Questionnaire. The Cogntion-FIM was found to demonstrate an excellent correlation with the Mini-Mental State Examination (MMSE); an adequate correlation with the Lowenstein Occupational Therapy Cognitive Assessment (LOTCA), Office of Population Censuses and Surveys Disability scores, and the revised Wechsler Adult Intelligence Test-verbal IQ; and a poor correlation with the London Handicap Scale, SF-36 Physical and Mental components, and the General Health Questionnaire.

Ecological:
The Motor-FIM demonstrated poor correlations with the Occupational Therapy Adult Perceptual Screening Test (OT-APST). The Cognition-FIM demonstrated adequate correlations with the OT-APST.

Does the tool detect change in patients? A significant ceiling effect has been detected with the Cognitive domain of the FIM. Out of seven studies examined, three reported that the FIM has an excellent ability to detect change in patients with stroke, four reported poor ability to detect change in patients with stroke or multiple sclerosis.
Acceptability The FIM is typically administered by interview. In patients with stroke, it can be reliability administered to proxy respondents.
Feasibility Training and education of persons to administer the FIM may represent significant cost. Use of interview formats may make the FIM more feasible for longitudinal assessment.
How to obtain the tool? Click here to find a copy of the FIM (the original comes from the following website: http://www.va.gov/vdl/documents/Clinical/Func_Indep_Meas/fim_user_manual.pdf)

http://www.udsmr.org

Psychometric Properties

Overview

We conducted a literature search to identify all relevant publications on the psychometric properties of the FIM.

Floor/Ceiling Effects

Van der Putten, Hobart, Freeman and Thompson (1999) compared the Motor-FIM and total FIM to the Barthel Index in 201 patients with multiple sclerosis and 82 post-stroke patients undergoing inpatient neurorehabilitation. The Cognitive-FIM had poor ceiling effects in patients with multiple sclerosis (36%) and adequate ceiling effects in patients with stroke. The total FIM showed no ceiling effect (0%) in both patients with stroke and patients with multiple sclerosis, as compared to 7% for the Barthel Index (1% for the Motor-FIM).

Hsueh, Lin, Jeng, and Hsieh (2002) compared the Motor-FIM, the original 10-item Barthel Index, and the 5-item short form Barthel Index in inpatients with stroke receiving rehabilitation. They reported a substantially larger floor effect for admission Barthel Index scores than for admission Motor-FIM scores (18.2% vs. 5.8% respectively).

Hobart and Thompson (2001) compared the modified Barthel Index, the FIM and the 30-item FIM plus Functional Assessment Measure (FIM + FAM) in 149 patients with various neurological disorders. No significant floor or ceiling effects were reported in this study for the total FIM, although there was a 16.1% ceiling effect noted for the Cognitive-FIM.

Brock, Goldie, and Greenwood (2002) examined the ceiling effects of the Motor-FIM and the Motor Assessment Scale in 106 rehabilitation inpatients with stroke at discharge. The ceiling effects for the Motor-FIM were adequate (16%), and 29% of the patients achieved the highest score on the hardest item of the Motor-FIM. In comparison, the Motor Assessment Scale had a ceiling effect of 25% (poor) and 35% of patients scored the highest score on the most difficult item.

Dromerick, Edwards, and Diringer (2003) assessed 95 consecutive admissions to a stroke rehabilitation service for disability on admission and discharge. No floor or ceiling effects were reported at admission to or discharge from rehabilitation with the FIM, whereas the Barthel Index demonstrated a large ceiling effect at discharge (27%).

Reliability

Internal consistency:
Dodds, Martin, Stolov and Deyo (1993) examined the psychometric properties of the FIM by analyzing Uniform Data System data on 11,102 general rehabilitation inpatients. Common diagnoses were stroke (52%), orthopedic conditions (10%), and brain injury (10%). The FIM demonstrated an excellent internal consistency, with a Cronbach’s alpha of 0.93 for overall admissions and 0.95 for discharges.

Hsueh, Lin, Jeng, and Hsieh (2002) examined the reliability of the FIM in 118 inpatients with stroke. Patients were administered the Motor-FIM subscale at admission to a rehabilitation ward of a hospital and before discharge from the hospital. The Motor-FIM demonstrated excellent internal consistency, with an alpha = 0.88 at admission and an alpha = 0.91 at discharge.

Hobart et al. (2001) examined the reliability of the FIM, the Barthel Index and the FIM plus Functional Assessment Measure in 149 rehabilitation inpatients with neurologic disorders. Item-to-total correlations were adequate and ranged from 0.53 to 0.87 for the FIM total, 0.60 for the Motor-FIM and 0.63 for the Cognitive-FIM. Mean inter-item correlations were also adequate, and were reported as 0.51 for the total FIM, 0.56 to 0.91 for the Motor-FIM and 0.72 to 0.80 for the Cognitive-FIM. Cronbach alpha levels were excellent for the total FIM (alpha = 0.95), the Motor-FIM (alpha = 0.95), and for the Cognitive-FIM (alpha = 0.89). The results of this study demonstrate the internal consistency of the total FIM and its Motor and Cognitive domains.

Sharrack, Hughes, Soudain, and Dunn (1999) assessed the internal consistency of the FIM in patients with multiple sclerosis. The internal consistency of the FIM was excellent, with a Cronbach’s alpha of 0.98.

Test-retest:
Chau, Daler, Andre and Patris (1994) examined the test-retest reliability of the FIM in 254 patients under 20 years old in a rehabilitation centre. The test-retest reliability was found to be excellent (ICC = 0.93 for total FIM).

Segal, Ditunno, and Staas (1993) examined the test-retest reliability of the FIM at discharge from an acute care rehabilitation setting and again at admission to an ongoing rehabilitation setting in 57 patients with spinal cord injuries. The two ratings were performed within 6 days of each other. The total FIM demonstrated excellent test-retest reliability (r = 0.83).

Kidd et al. (1995) compared the FIM to the Barthel Index in two groups of 25 patients undergoing neurorehabilitation. Test-retest reliability was found to be excellent for the FIM (r = 0.90).

Ottenbacher, Hsu, Granger, and Fiedler (1996) examined the test-retest reliability of the FIM by examining the results of 11 studies including a total of 1,568 patients. The median test-retest was excellent (r = 0.95).

Pollak, Rheault, and Stoecker (1996) assessed the test-retest reliability of the FIM in 49 individuals over the age of 80 years. Individuals were tested twice using the FIM. Excellent test-retest reliability was found for the Motor-FIM (ICC = 0.90), and for the Cognitive-FIM (ICC = 0.80).

Intra-rater:
Sharrack, Hughes, Soudain, and Dunn (1999) assessed the intra-rater reliability of the FIM (using both kappa and ICC statistics) in 35 patients with multiple sclerosis. Three raters followed patients for 9 months, with assessments every 3 months. The kappa value for the total FIM was poor (kappa = 0.28), however the ICC was excellent (ICC = 0.94). For individual items, kappa coefficients ranged from adequate (kappa = 0.55 for Dressing-lower body) to excellent (kappa = 1.00 for both Expression and Social interaction). ICC’s for individual items ranged from adequate (kappa = 0.60 for Bladder control) to excellent (ICC = 1.00 for both Expression and Social interaction).

Hobart et al. (2001) examined the intra-rater reliability of the FIM, the Barthel Index and the FIM plus Functional Assessment Measure in 56 rehabilitation inpatients with neurologic disorders. Patients were examined by the same multidisciplinary team on two occasions. Intra-rater reliability was calculated using ICC statistics. The total FIM, Motor-FIM and Cognitive-FIM were all found to have excellent intra-rater reliabilities (ICC = 0.98, 0.98 and 0.95, respectively).

Inter-rater:
Chau, Daler, Andre and Patris (1994) examined the inter-rater reliability of the FIM between educators, occupational therapists and physiotherapists in 254 patients under 20 years old in a rehabilitation centre. Inter-rater reliability for the total FIM was excellent (ICC = 0.94).

Ottenbacher, Mann, Granger, Tomita, Hurren, and Charvat (1994) examined the inter-rater reliability of the FIM and the Instrumental Activities of Daily Living Scale in 20 community-dwelling older patients. Two raters administered the tests over a short (7-10 days) or long (4-6 week) interval. The ICCs for inter-rater reliability were excellent, ranging from 0.90 to 0.99.

Ottenbacher, Hsu, Granger, and Fiedler (1996) examined the inter-rater reliability of the FIM by examining the results of 11 studies including a total of 1,568 patients. The median inter-rater reliability for the total FIM was excellent (r = 0.95).

Hamilton, Laughlin, Fiedler and Granger (1994) examined the inter-rater reliability of the FIM in 1,018 patients. The total FIM ICC was excellent (ICC = 0.96), as was the Motor-FIM domain (ICC = 0.96), and the Cognitive-FIM domain (ICC = 0.91).

Jaworski, Kult, and Boynton (1994) compared the reliability of observed and reported FIM ratings. In this study, the inter-rater reliability of the FIM was found to be excellent (ICC = 0.99).

Kidd et al. (1995) compared the FIM to the Barthel Index in two groups of 25 patients undergoing neurorehabilitation. Inter-rater reliability was found to be excellent for the FIM (r = 0.92).

Segal and Schall (1994) examined the inter-rater reliability of the FIM in 38 patients with stroke. The inter-rater reliability of the measure was found to be excellent, with an ICC of 0.96.

Brosseau and Wolfson (1994) examined the inter-rater reliability of the FIM in patients with multiple sclerosis and found that the FIM has an excellent inter-rater reliability (ICC = 0.83).

Daving, Andren, Nordholm, and Grimby (2001) examined the reliability of the FIM in 63 patients with stroke, approximately 2 years after stroke onset. Two raters (between three occupational therapists and one nurse) conducted independent ratings of the FIM in the patient’s home, and the interview procedure was repeated within a week by another two raters in the clinic. The kappa values during the same interview exceeded 0.40 for 17 items, demonstrating adequate to excellent inter-rater reliability , however, the Social interaction item kappa value was poor (kappa = 0.26). In comparing the two interviews, kappa values were between 0.40-0.60 for Self-care items (except Bathing) and Sphincter control (except Bowel management), however, most of the Transfers, Locomotion and Social cogniton items had kappa values below 0.40. The two interviews were also studied using ICC statistics between all raters. ICC’s ranged from adequate (0.62 for Bowel management) to excellent (0.88 for Bathing) for the 13 motor items, and were adequate (ranging from 0.60 to 0.72) for the Cognitive domain, except for the Social interaction item which had an ICC of only 0.44. Significant differences were found between raters on the Wilcoxon test for the Dressing, Transfer Toilet, Transfer Tub/Shower, Walk/Wheelchair and the Cognitive domain. The results of this study show that the FIM demonstrates high inter-rater reliability in the same interview setting (whether at home or at the clinic), however the stability over time with a repeated interview by different raters is less reliable.

Sharrack, Hughes, Soudain, and Dunn (1999) assessed the inter-rater reliability of the FIM (using both kappa and ICC statistics) in 64 patients with multiple sclerosis. Each patient was assessed by three raters (2 neurologists, 1 neurology research nurse). The kappa value for the total FIM score was poor (kappa = 0.21), however the ICC was excellent (ICC = 0.99). For individual items, kappa coefficients were variable and ranged from poor (kappa = 0.26 for Comprehension) to excellent (kappa = 0.88 for Stairs locomotion). ICC’s for the individual items were excellent, ranging from 0.76 to 0.99 with the exception of the Comprehension item, which demonstrated adequate inter-rater reliability (ICC = 0.56).

Validity

Content:

The FIM was created based on the results of a literature review of published and unpublished measures and expert panels. To establish content and face validity, the FIM was then piloted in 11 centers (including 114 clinicians from 8 different disciplines and 110 patients evaluated) (Keith & Granger, 1987). Face and content validity were both determined by applying the Delphi method, using rehabilitation expert opinion to establish the inclusiveness and appropriateness of the items (Granger, Hamilton, Keith, Zielezny, & Sherwins, 1986).

Criterion:

Concurrent:
Hsueh, Lin, Jeng, and Hsieh (2002) examined the concurrent validity of the Motor-FIM by examining its interrelations with the original 10-item Barthel Index, and the 5-item short form Barthel Index in 118 inpatients with stroke receiving rehabilitation. Concurrent validity was measured using ICC and Spearman correlations. The Motor-FIM exhibited excellent concurrent validity at admission as measured by Spearman correlation (r = 0.74) and adequate validity as measured by ICC (ICC = 0.55). The Motor-FIM exhibited excellent concurrent validity at discharge (Spearman correlation = 0.92, ICC = 0.86).

Kwon, Hartzema, Duncan and Min-Lai (2004) examined the concurrent validity of the Barthel Index, the FIM and the Modified Rankin Scale in a sample of post-stroke patients. Spearman correlation coefficients were excellent between the Barthel Index and the Motor-FIM (r = 0.95) and between the Motor-FIM and the Modified Rankin Scale (r = -0.89).
Note: This correlation is negative because a high score on the FIM indicates functional independence, whereas a high score on the Modified Rankin Scale indicates severe disability).

Hall, Hamilton, Gordon, and Zasler (1993) examined the concurrent validity of the Disability Rating Scale, the FIM, and the Functional Assessment Measure. Excellent correlations were found between the Motor-FIM and Cognition-FIM and the Disability Rating Scale (r = 0.64 and 0.73, respectively).

Zwecker et al. (2002) examined the relationship between cognitive status and functional motor outcomes in 66 patients with stroke. Functional motor outcomes were measured from efficacy and efficiency of the FIM motor scores (isolated from total FIM scores) and the Montebello Rehabilitation Factor Score (MRFS). Using Pearson’s correlation, an adequate correlation was found between the FIM cognitive subtest and MRFS efficacy (r=0.34, p<0.01). A poor correlation was found between the FIM cognitive and MRFS efficiency (r=0.28, p<0.05). No significant correlations were found between the FIM cognitive and FIM motor efficacy or efficiency scores.

Predictive:
For an extensive review of the predictive validity of the FIM, please see:

Timbeck, R. J., Spaulding, S. J. (2003). Ability of the Functional Independence Measure to predict rehabilitation outcomes after stroke: A review of the literature. Physical & Occupational Therapy in Geriatrics, 22(1), 63-76.

Chumney, D., Nollinger, K., Shesko, K., Skop, K., Spencer, M., Newton, R.A. (2010). Ability of Functional Independent Measure to accurately predict functional outcome of stroke-specific population: Systematic review. Journal of Rehabilitation and Development, 47, 17-30.

Predictive validity of the FIM in the amount of care patients require in their homes:
Granger, Cotter, Hamilton, Fiedler, and Hens (1990) examined whether the FIM could predict the amount of help (measured in minutes of assistance provided per day by another person in the home) patients with multiple sclerosis required, using a bivariate regression analysis. Burden of care was assessed as help in minutes per day. It was found that a 1-point improvement in total FIM score predicted a 3.38-minutes reduction in help from another person per day. The FIM was found to be more predictive than the Barthel Index, the Incapacity Status Scale, and the Environmental Status Scale. The FIM was also found to contribute to the prediction of patient general life satisfaction.

Granger, Cotter, Hamilton and Fiedler (1993) examined whether the FIM could predict the physical care needs (measured in minutes of assistance provided per day by another person in the home) of patients with stroke. Burden of care was assessed as help in minutes per day. It was found that a 1-point improvement in total FIM score predicted a 2.19-minute reduction in help from another person per day. The FIM, along with the Brief Symptom Inventory, was found to contribute to the prediction of patient general life satisfaction.

Corrigan, Smith-Knapp and Granger (1997) examined the predictive validity of the FIM for patients with traumatic brain injury after discharge from acute rehabilitation. They found that the Motor-FIM predicted which patients required direct assistance with 83% accuracy, the Cognitive-FIM predicted which patients required supervision with 77% accuracy, and the Motor-FIM and Cognitive-FIM predicted which patients required any assistance with 78% accuracy. Further, the Motor-FIM score alone was the best predictor of the number of minutes of assistance needed.

Predictive validity of the FIM with discharge FIM scores, discharge destinations, length of stay, functional gain, depression, survival, and the ability to return to work following stroke or traumatic brain injury:
Inouye et al. (2000) performed a multivariate analysis on data from rehabilitation patients with stroke obtained from patient medical records to identify predictors of functional outcome using total FIM scores. It was found that total FIM admission scores was the strongest predictor of total FIM discharge scores. No relationship was found between total FIM scores at discharge and gender, hospital length of stay, or the nature of the stroke.

Oczkowski and Barreca (1993) examined whether the FIM could predict prognosis in 113 patients with stroke observed from admission to discharge. It was found that the admission FIM score was predictive of placement after discharge and of outcome disability. No patients with an admission FIM score below 36 were discharged home, while all of the patients with admission FIM scores above 96 were discharged home. However, discharge destination became difficult to predict in patients with a moderate range of disability (i.e. an FIM score > 36 or < 97). When individual FIM items were considered, a patient’s level of independence with bowel and bladder management was predictive of functional outcome and discharge destination.

Alexander (1994) examined the predictive validity of the FIM in a sample of 520 patients with stroke admitted to a rehabilitation hospital. It was found that an admission FIM score of < 40 resulted in an acute care stay almost twice as long as any other FIM score. Patients aged < 55 years all were discharged home regardless of their initial severity. Patients with an FIM score < 40 and who were > 55 years old had a 50% chance of being discharged to a long term care facility. This is in contrast to the findings by Oczkowski and Barreca (1993) who found that no patients with an admission FIM score < 36 were discharged to home. Patients with an admission FIM score between 40-60 who were > 74 years were at high risk for discharge to a long term care facility. Patients with an FIM score > 80 were discharged home.

Mokler, Sandstrom, Griffin, Farris, and Jones (2000) found that in the acute care phase of stroke recovery, the FIM scores for Eating, Bathing, Dressing – Lower body, Toileting, Bowel management and Social interaction and predicted discharge destination with 70% accuracy. In the later phase of recovery in rehabilitation, particularly in patients with a severe stroke, scores on admission FIM items including Bladder management, Toilet transfer, and Memory, and scores on the discharge FIM items including Dressing – Upper body, Bed/Chair/Wheelchair transfers and Comprehension were associated with predicting discharge destination with up to 75% accuracy. These three admission items and three discharge items correctly predicted discharge placement in 2/3 and 3/4 of the cases, respectively.

Black, Soltis, and Bartlett (1999) examined the FIM scores of 234 patients with stroke admitted to a rehabilitation facility over a 2-year period. Patients who were discharged home were less likely to have a caregiver who worked (20%) versus patients who were discharged to long-term care (65%). The availability of a non-working family member to provide assistance and supervision was a critical factor related to discharge home. Patients with a discharge FIM score > 80 had a high probability of being discharged home when social factors (e.g. availability of family support and non-working family member) were taken into consideration. Thus, both functional status and social factors, such as family availability and support, are critical elements in predicting the discharge destination of this patient population.

Ring et al. (1997) examined 151 patients with stroke admitted to a rehabilitation centre over a 2-year period. They found that admission FIM scores and length of stay were the most significant predictors of functional gain.

Heinemann, Linacre, Wright, Hamilton, and Granger (1994) examined the extent to which functional outcome measures could predict functional status in patients with traumatic brain injury. They report that admission FIM scores were related to discharge function and length of stay. Admission Motor-FIM scores were found to be a stronger predictor of length of stay than Cognitive-FIM scores and accounted for 52% of the variance in discharge motor function. Admission Cognitive-FIM scores accounted for 46% of the variance in discharge cognitive function.

Ween, Mernoff, and Alexander (2000) examined the predictive validity of the FIM in 244 patients with stroke at an acute rehabilitation centre. It was found that patients with an admission FIM score < 50 were dependent in their self-care activities upon discharge. Patients who scored < 70, nine days post-stroke, were highly likely to remain functionally dependent at discharge. Patients who scored > 70 were not dependent at discharge and had shorter than average length of stay. Patients who scored between 50 and 70 on the admission FIM had unpredictable outcomes. In terms of discharge destination, patients who were < 60 years old and had an admission FIM score > 70 were strongly associated with home discharge.

Stineman, Fiedler, Granger, and Maislin (1998) examined the records of 26,339 patients with stroke discharged from 252 inpatient rehabilitation facilities. They found that patients whose admission FIM scores were > 37 were able to eat, groom, dress their upper bodies and manage their bowels and bladder independently at discharge. Patients who scored > 55 were also able to bathe, dress their lower bodies and transfer onto a bed or chair and toilet. Additionally, most patients who had initial Motor-FIM scores > 62 and whose Cognitive-FIM scores were > 30 gained independence in most tasks, including transferring into the tub and climbing the stairs by the time of discharge. They also found that between 85% and 93% of patients with moderate stroke were discharged home.

Singh et al. (2000) administered the FIM to 81 patients with stroke at 1 month, 3 months, and 1 year post-stroke. Using stepwise linear regression, they found that lower total FIM scores at 1-month post-stroke were predictive of higher depression scores at 3 months post-stroke.

Cifu et al. (1997) compared 49 patients with traumatic brain injury who were employed at one-year follow-up with 83 patients who remained unemployed at one-year. They found that FIM scores at admission to rehabilitation were significantly associated with patients’ employment status one-year post head injury, such that patients who had returned to work one-year later had demonstrated significantly higher scores on the FIM at admission.

Tur, Gursel, Yavuzer, Kucukdeveci and Arasil (2003) examined the predictive validity of the FIM in 102 patients with stroke admitted to rehabilitation units. The FIM was administered within 72 hours of admission and at discharge. Using a stepwise regression analysis, FIM scores at admission were found to be excellent predictors of FIM scores at discharge (0.90; p<0.001), indicating that the FIM can be used to predict functional recovery in patients with stroke.

Whiting, Shen, Hung, Cordato & Chan (2010) examined predictors of 5-year survival in 166 patients with stroke (mean age 80 years), using the FIM. Using a logistic regression model, lower preadmission FIM scores were found to negatively predict 5-year survival of patients with stroke (OR 1.04, 95%CI 1.1-2.0, P=0.01). In addition, total FIM scores were found to remain relatively stable from baseline to 5-year follow up in the 5-year survival group, however, FIM cognition scores were lower than baseline scores at the 5-year follow-up.

Predictive validity of the FIM in patients with aphasia and neglect:
Granger, Hamilton, and Fielder (1992) found that at admission and discharge, functional scores for patients with right brain damage were slightly higher, but length of stay at the hospital and rate of community discharge were similar to that of patients who had left brain damage.

Alexander (1994) found that patients with stroke who had severe right brain damage had significantly less FIM change than patients with severe left brain damage.

Ring et al. (1997) found that patients with neglect or aphasia had significantly higher FIM gains despite lower FIM admission scores. However, these patients also had a much longer length of stay at the hospital. It was also found that 96% of patients with right brain damage without neglect and 88% of patients with right brain damage and neglect were discharged home.

Oczkowski and Barreca (1993) found that patients with any degree of hemianopsia, parietal neglect, aphasia, or cognitive impairment had significantly lower FIM scores than those without these impairments, but unlike the results of Ring et al. (1997), hemianopsia, side of lesion, neglect and aphasia were not predictive of discharge destination.

Katz et al. (2000) examined correlations between the FIM (total, motor and cognitive scores) and the Lowenstein Occupational Therapy Cognitive Assessment (LOTCA – Orientation, Perception, Visuomotor Organisation and Thinking Operations subtest) in two subgroups of adults with right hemisphere stroke (n=40 vs. patients without unilateral spatial neglect, n=21), using Spearman’s correlation analysis. Measures were taken on admission to and discharge from rehabilitation, and at 6-month follow-up. In the neglect group, adequate correlations were reported between FIM total and FIM motor, and LOTCA Visuomotor Organisation and Thinking Operations (range r=0.48 to -.51) at admission. Adequate to excellent correlations were reported between FIM total and FIM motor, and LOTCA Perception, Visuomotor Organisation and Thinking Operations (range r=0.48 to 0.75) at discharge. Excellent correlations were reported between FIM total and FIM motor and LOTCA Visuomotor Organisation and Thinking Operations tasks (range r=0.61 – 0.77) at follow-up. In the non-neglect group, poor to excellent correlations were reported between FIM cognitive and LOTCA scores (range r=0.05 to -.67) at admission. Moderate to excellent correlations were reported between FIM total and FIM motor, and LOTCA Visuomotor Organisation and Thinking Operations tasks at discharge and follow-up (range r=0.43 to 0.62).
Note: The FIM cognitive was not readministered at discharge or follow-up with this subgroup.

Construct:

Linacre et al. (1994) applied Rasch analysis to the admission and discharge FIM scores of 14,799 patients. Two distinct aspects of disability were found within the FIM: Motor and Cognitive function.

Cavanagh, Hogan, Gordon, and Fairfax (2000) suggested that for post-stroke patients, a simple 2-factor model of the FIM may be insufficient to describe disability and may not measure within patient change adequately. The authors suggest that a three-dimensional FIM for patients with stroke be applied, which includes Self-care, Cognitive function, and Toileting as the major grouping of scales. They found that the 2-factor model only accounts for 66% of variance, whereas a 3-factor model accounted for more variance (74.2%).

Convergent/Discriminant:
Hobart et al. (2001) found that the total FIM and Motor-FIM scores correlated more strongly with the Office of Population Censuses and Surveys Disability Scales disability scores (r = 0.82 and 0.84, respectively), London Handicap Scale scores (r = 0.32 and 0.35, respectively), the SF-36 Physical component scores (r = 0.26 and 0.30, respectively) and the revised Wechsler Adult Intelligence Test-verbal IQ test (r = 0.35 and 0.27, respectively), than with measures of mental health status (SF-36 Mental component, r = 0.10 and 0.10, respectively) or psychological distress (General Health Questionnaire, r = 0.13 and r = 0.15, respectively). However, the Cognitive-FIM correlated most strongly with Office of Population Censuses and Surveys Disability scores (r = 0.43) and the revised Wechsler Adult Intelligence Test-verbal IQ scores (r = 0.51) and correlated poorly with the London Handicap Scale (r = 0.11), the SF-36 Physical and Mental components (r = 0.04 and r = 0.08, respectively), and the General Health Questionnaire (r = 0.01).

Giaquinto, Giachetti, Spiridigliozzi and Nolfe (2010) examined the convergent validity of the FIM, Hospital Anxiety Depression Scale (HADS) and the World Health Organization Quality of Life scale (WHOQOL-100) in 107 patients with stroke (mean 5.6 months post-stroke). Assessments were performed at admission and discharge from a two-month rehabilitation program. As measured by Pearson’s correlation coefficients, an excellent correlation was found between FIM admission and FIM discharge scores (r=0.656, p<0.0001) and was not significantly influenced by gender. However, correlations between FIM discharge scores and HADS and WHOQOL-100 scores were influenced by gender. Among females an adequate correlation was found between FIM discharge and HADS scores (r=-0.315, p<0.02) and FIM discharge and WHOQOL-100 scores (r=0.339, p<0.01), but the correlations among males’ scores were poor (r=0.139 and r=0.147 respectively).

Zwecker et al. (2002) reported an adequate correlation between the FIM cognitive subtest and the Lowenstein Occupational Therapy Cognitive Assessment (LOTCA) (r= 0.471, p<0.001) and an excellent correlation between the FIM cognitive subtest and the Mini Mental State Examination (MMSE) (r=0.666, p<0.001) in 66 patients with stroke, using Pearson’s Correlation.

Known groups:
Dodds, Martin, Stolov and Deyo (1993) examined the construct validity of the FIM using data from 11,102 general rehabilitation inpatients (52% with stroke, 10% with orthopedic conditions, 10% with brain injury). FIM scores discriminated between groups based on spinal cord injury and stroke severity, and the presence of comorbid illness both at admission and discharge. The communication item of the FIM demonstrated most of the observed score difference.

Ring, Feder, Schwartz, and Samuels (1997) examined 151 patients with stroke admitted to a rehabilitation centre over a 2-year period. They found that the FIM was able to distinguish between patients with or without neglect and with or without aphasia at both admission and discharge.

Ecological validity:

Cooke, McKenna, Fleming & Darnell (2006) examined the ecological validity of the Occupational Therapy Adult Perceptual Screening Test (OT-APST) by comparing scores and completion time with the FIM motor and cognitive subtests in a sample of patients with stroke (n=208). Significant but poor correlations were reported between FIM motor scores and 6 of the 7 OT-APST subscales (range r=0.26 to 0.41, p<0.01). Significant adequate correlations were reported between FIM cognitive scores and all 7 OT-APST subscales (range r=0.36 to 0.50, p<0.01). Significant poor to adequate negative correlations were also reported between the time taken to complete the FIM motor and cognitive subtests and the OT-APST (r=-0.27 and -0.33 respectively, p<0.01).

Responsiveness

The FIM is often compared to the Barthel Index, because the FIM was developed to be a more comprehensive and responsive measure of disability than the Barthel Index (van der Putten et al., 1999; Hobart & Thompson, 2001; Wallace, Duncan, & Lai, 2002; Hsueh et al., 2002).

Van der Putten et al. (1999) compared the Motor-FIM and total FIM to the Barthel Index in 201 patients with multiple sclerosis and 82 post-stroke patients undergoing inpatient neurorehabilitation. The Motor-FIM and total FIM demonstrated small effect sizes in the expected direction from admission to discharge in patients with multiple sclerosis (ES = 0.34 and ES = 0.30, respectively) and large effect sizes in patients with stroke (ES = 0.91 and ES = 0.82). The effect sizes for the Cognitive-FIM were not significant (ES = 0) in patients with multiple sclerosis and moderate in patients with stroke (ES = 0.61). Change scores for all scales in both disease groups were positive, indicating less disability on discharge than admission. Effect sizes on the Barthel Index were similar to those of the FIM in both patient groups, suggesting that the FIM might not have an advantage in terms of its responsiveness to change.

Wallace et al. (2002) compared the responsiveness of the Motor-FIM to the Barthel Index for stroke recovery between 1 and 3 months. The Barthel Index and Motor-FIM exhibited similar responsiveness to change in this patient population (Motor-FIM, ES = 0.28; Standardized Response Mean (SRM) = 0.62; AUC/ROC curve = 0.675).

Hsueh et al. (2002) compared the responsiveness of the Motor-FIM, the original 10-item Barthel Index, and the 5-item short form Barthel Index in inpatients with stroke receiving rehabilitation. The Barthel Index and Motor-FIM exhibited high responsiveness (SRM = 1.2), indicating significant change.

Dromerick et al. (2003) assessed 95 consecutive admissions to a stroke rehabilitation service for disability on admission and discharge. The Modified Rankin Scale and the International Stroke Trial Measure were compared with the Barthel Index and the FIM. The number of patients for which each scale detected a clinically significant change in disability was determined. The SRM of the FIM was superior to that of the Barthel Index (2.18 versus 1.72) (change from admission to discharge from rehabilitation). The FIM was the most sensitive measure, detecting change in 91/95 subjects, including change in 18 patients in whom the Barthel Index detected no change.

Hobart and Thompson (2001) compared the responsiveness of the modified Barthel Index, the FIM and the 30-item FIM plus Functional Assessment Measure (FIM + FAM) in 149 patients with various neurological disorders. The SRMs for the Barthel Index, the FIM, and the FIM + FAM scales measuring global, motor, and cognitive disability were found to be similar, suggesting that there is no advantage in responsiveness of one measure over another (total FIM, SRM = 0.48; Motor-FIM, SRM = 0.54; Cognitive-FIM, SRM = 0.17).

Sharrack et al. (1999) examined the responsiveness of the FIM in 25 patients with multiple sclerosis. Patients were followed for 9 months, with assessments every 3 months. The total FIM demonstrated a poor sensitivity to change (ES = 0.46). A number of motor items (i.e. Eating, Grooming, Sphincter control, Bed/Chair/Wheelchair and Toilet Transfers, and Locomotion) had small to moderate responsiveness (ES ranged from 0.25 for Toilet Transfer to 0.67 for Bed/Chair/Wheelchair Transfers). None of the cognitive items were responsive to change (ES ranged from 0.00 to 0.19).

Dodds, Martin, Stolov and Deyo (1993) examined the responsiveness of the FIM by analyzing the differences between admission and discharge FIM scores from 11,102 general rehabilitation inpatients (with stroke (52%), orthopedic conditions (10%), and brain injury (10%)). Significant functional gains were detected by the FIM (33% score improvement). The authors conclude that the FIM demonstrates some responsiveness, but its ability to measure change over time needs further examination.

Hammond, Grattan, Sasser, Corrigan, Bushnik, and Zafonte (2001) examined FIM score changes over time in patients with traumatic brain injury. Significant differences in total FIM, Motor-FIM and Cognitive-FIM scores were reported between discharge from rehabilitation and follow-up at one year post-injury. Change between one and two years and one and five years was reported to be distributed across all items with most change observed in cognitive function.

Beninato, Gill-Body, Salles, Stark, Black-Schaffer and Stein (2006) defined the minimal clinically important difference (MCID) when using the FIM in a stroke population. The study included 113 patients from a rehabilitation unit at a long-term acute care hospital. The FIM was administered at admission and discharge; patient function was also assessed by attending physicians at the same time points using a 15-point integer scale where -7 indicated that a patient was “a very great deal worse”, 0 indicated “no change” and +7 indicated “a very great deal better”. Based on physicians’ ratings of clinical change made at discharge, change scores of 22, 17 and 3 for total FIM, motor FIM and cognitive FIM (respectively), were deemed to differentiate patients who demonstrated clinically important change from those who had not. Generalization of results is cautioned as the study only included patients receiving treatment at one centre and patient, caregiver or family assessments were not included in the ratings of important change.

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See the measure

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Table of contents

Multiple Errands Test (MET)

Evidence Reviewed as of before: 08-05-2013
Author(s)*: Valérie Poulin, OT, PhD candidate; Annabel McDermott, OT
Editor(s): Nicol Korner-Bitensky, PhD OT
Expert Reviewer: Deirdre Dawson, PhD OT

Purpose

The Multiple Errands Test (MET) evaluates the effect of executive function deficits on everyday functioning through a number of real-world tasks (e.g. purchasing specific items, collecting and writing down specific information, arriving at a stated location). Tasks are performed in a hospital or community setting within the constraints of specified rules. The participant is observed performing the test and the number and type of errors (e.g. rule breaks, omissions) are recorded.

In-Depth Review

Purpose of the measure

The Multiple Errands Test (MET) evaluates the effect of executive function deficits on everyday functioning through a number of real-world tasks (e.g. purchasing specific items, collecting and writing down specific information, arriving at a stated location). Tasks are performed in a hospital or community setting within the constraints of specified rules. The participant is observed performing the test and the number and type of errors (e.g. rule breaks, omissions) are recorded.

The Multiple Errands Test was developed by Shallice and Burgess in 1991. The measure was intended to evaluate a patient’s ability to organize performance of a number of simple unstructured tasks while following several simple rules.

See Alternative Forms sections below for information regarding other versions.

Features of the measure

Items:

The original Multiple Errands Test (Shallice and Burgess, 1991) was comprised of 8 items: 6 simple tasks (e.g. buy a brown loaf of bread, buy a packet of throat pastilles), 1 task that is time-dependent, and 1 that comprises 4 subtasks (see Description of tasks, below). It should be noted that the MET was originally devised in an experimental context, rather than as a formal assessment.

Description of tasks:

The original Multiple Errands Test (Shallice and Burgess, 1991) was comprised of 8 written tasks to be completed in a pedestrian shopping precinct. Tasks and rules are written on a card provided to the participant before arriving at the shopping precinct. Of the 8 tasks, 6 are simple (e.g. buy a brown loaf of bread, buy a packet of throat pastilles), the 7th requires the participant to be at a particular place 15 minutes after starting the test, and the 8th is more demanding as it comprises 4 sets of information that the participant must obtain and write on a postcard:

  1. the name of the shop most likely to have the most expensive item;
  2. the price of a pound of tomatoes;
  3. the name of the coldest place in Britain yesterday; and
  4. the rate of the exchange of the French franc yesterday.

The card also includes instructions and rules, which are repeated to the participant on arrival at the shopping precinct:

“You are to spend as little money as possible (within reason) and take as little time as possible (without rushing excessively). No shop should be entered other than to buy something. Please tell one or other of us when you leave a shop what you have bought. You are not to use anything not bought on the street (other than a watch) to assist you. You may do the tasks in any order.“

Scoring:

The participant is observed performing the test and errors are recorded according to the following categorizations:

  • Inefficiencies: where a more effective strategy could have been applied
  • Rule breaks: where a specific rule (either social or explicitly mentioned in the task) is broken
  • Interpretation failure: where requirements of a particular task are misunderstood
  • Task failure: where a task is either not carried out or not completed satisfactorily.

Time taken to complete the assessment is recorded and the total number of errors is calculated.

Alternative versions of the Multiple Errands Test

Different versions of the MET were developed for use in specific hospitals (MET – Hospital Version and Baycrest MET), a small shopping plaza (MET – Simplified Version), and a virtual reality environment (Virtual MET). For each of these versions, 12 tasks must be performed (e.g. purchasing specific items and collecting specific information) while following several rules.

MET – Hospital Version (MET-HV – Knight, Alderman & Burgess, 2002)

The MET-HV was developed for use with a wider range of participants than the original version by adopting more concrete rules and simpler tasks. Clients are provided with an instruction sheet that explicitly directs them to record designated information. Clients must achieve four sets of simple tasks, with a total of 12 separate subtasks:

  1. The client must complete six specific errands (purchase 3 items, use the internal phone, collect an envelope from reception, and send a letter to an external address).
  2. The client must obtain and write down four items of designated information (e.g. the opening time of a shop on Saturday).
  3. The client must meet the assessor outside the hospital reception 20 minutes after the test had begun and state the time.
  4. The client must inform the assessor when he/she finishes the test.

The MET-HV uses 9 rules in order to reduce ambiguity and simplify task demands (Knight et al., 2002). Errors are categorized according to the same definitions as the original MET. The test is preceded by (a) an efficiency question rated using an end-point weighted 10-point Likert scale (“How efficient would you say you were with tasks like shopping, finding out information, and meeting people on time?“); and (b) a familiarity question rated using a 4-point scale (“How well would you say you know the hospital grounds?“). On completion the client answers a question rated using a 10-point scale (“How well do you think you did with the task?“).

MET – Simplified Version (MET-SV – Alderman, Burgess, Knight & Henman, 2003)

The MET-SV includes four sets of simple tasks analogous to those in the original MET, however the MET-SV incorporates 3 main modifications to the original version:

  1. More concrete rules to enhance task clarity and reduce likelihood of interpretation failures;
  2. Simplification of task demands; and
  3. Space provided on the instruction sheet for the participant to record the information they were required to collect.

The MET-SV has 9 rules that are more explicit than the original MET and are clearly presented on the instruction sheet.

Baycrest MET (BMET – Dawson, Anderson, Burgess, Cooper, Krpan & Stuss, 2009)

The BMET was developed with an identical structure to the MET-HV, except that some items, information and a meeting place are specific to the testing environment (Baycrest Center, Toronto). The BMET comprises 12 items and 8 rules. The test manual provides explicit instructions including collecting test materials, language to be used in describing the test, and a pretest section to ensure participants understand the tasks. Scoring was standardized to allow for increased usability. The score sheet allows identification of specific task errors or omissions, other inefficiencies, rule breaks and strategy use (please contact the authors for further details regarding the manual: ddawson@research.baycrest.org).

Virtual MET (VMET – Rand, Rukan, Weiss & Katz, 2009)

The VMET was developed within the Virtual Mall, a functional video-capture virtual shopping environment that consists of a large supermarket with 9 aisles. The system includes a single camera that films the user and displays his/her image within the virtual environment. The VMET is a complex shopping task that includes the same number of tasks (items to be bought and information to be obtained) as the MET-HV. However, the client is required to check the contents of the shopping cart at a particular time instead of meeting the tester at a certain time. Virtual reality enables the assessor to objectively measure the client’s behaviour in a safe, controlled and ecologically valid environment. It enables repeated learning trials and adaptability of the environment and task according to the client’s needs.

What to consider before beginning:

The MET is performed in a real-world shopping area that allows for minor unpredicted events to occur.

Time:

The BMET takes approximately 60 minutes to administer (Dawson et al., 2009).

Training requirements:

It is advised that the assessor reads the test manual and becomes familiar with the procedures for test administration and scoring.

Equipment:

  • Access to a shopping precinct or virtual shopping environment
  • Pen and paper
  • Instruction sheet (according to version being used)

Client suitability

Can be used with:

  • The MET has been tested on populations with acquired brain injury including stroke.

Should not be used with:

  • The MET cannot be administered to patients who are confined to bed.
  • Participants require sufficient language skills.
  • Some tasks may need to be adapted depending on the rehabilitation setting.

In what languages is the measure available?

The MET was developed in English.

Summary

What does the tool measure? The effect of executive function deficits on everyday functioning.
What types of clients can the tool be used for? The Multiple Errands Test can be used with, but is not limited to, clients with stroke.
Is this a screening or assessment tool? Assessment
Time to administer Baycrest MET: approximately 60 minutes (Dawson et al., 2009).
Versions
  • Multiple Errands Test (MET) (Shallice and Burgess, 1991)
  • MET – Simplified Version (MET-SV) (Alderman et al., 2003)
  • MET – Hospital Version (MET-HV) (Knight, Alderman & Burgess, 2002)
  • Virtual MET (Rand, Rukan, Weiss & Katz, 2009)
  • Baycrest MET (Dawson et al., 2009)
  • Modified version of the MET-SV and MET-HV (including 3 alternate versions) (Novakovic-Agopian et al., 2011, 2012)
Other Languages N/A
Measurement Properties
Reliability Internal consistency:
One study reported adequate internal consistency of the MET-HV in a sample of patients with chronic acquired brain injury including stroke.

Test-retest:
No studies have reported on the test-retest reliability of the MET with a population of patients with stroke.

Intra-rater:
No studies have reported on the intra-rater reliability of the MET with a population of patients with stroke.

Inter-rater:
– One study reported excellent inter-rater reliability of the MET-HV in a sample of patients with chronic acquired brain injury including stroke.
– One study reported adequate to excellent inter-rater reliability of the BMET in a sample of patients with acquired brain injury including stroke.

Validity Criterion:
Concurrent:
No studies have reported on the concurrent validity of the MET in a stroke population.

Predictive:
One study examined predictive validity of the MET-HV with a sample of patients with acquired brain injury including stroke and reported poor to adequate correlations between discharge MET-HV performance and community participation measured by the Mayo-Portland Adaptability Inventory (MPAI-4).

Construct:
Convergent/Discriminant:
– Three studies* examined convergent validity of the MET-HV and reported excellent correlations with the Modified Wisconsin Card Sorting Test (MWCST), Behavioural Assessment of Dysexecutive Syndrome battery (BADS), Dysexecutive questionnaire (DEX), IADL questionnaire and FIM Cognitive score; and an adequate correlation with the Rivermead Behavioural Memory Test (RBMT).
– One study* examined convergent validity of the MET-SV and reported adequate correlations with the Weschler Adult Intelligence Scale – Revised Full Scale IQ (WAIS-R FSIQ), MWCST, BADS and Cognitive Estimates test; and poor to adequate correlations with the DEX.
– One study* examined convergent validity of the BMET and reported adequate to excellent correlations with the Sickness Impact Profile and Assessment of Motor and Process Skills.
– Three studies* examined convergent validity of the VMET and reported excellent correlations with the MET-HV, BADS, IADL questionnaire, Semantic Fluencies test, Tower of London test, Trail Making Test, Corsi’s supra-span test, Street’s Completion Test and the Test of Attentional Performance.
*Note: Correlations between the MET and other measures of everyday executive functioning and IADLs used in these studies also provide support for the ecological validity of the MET.

Known Groups:
– Two studies reported that the MET-HV is able to differentiate between individuals with acquired brain injury (including stroke) vs. healthy adults, and between healthy older adults vs. healthy younger adults.
– One study reported that the MET-SV is able to differentiate between clients with brain injury including stroke vs. healthy adults.
– One study reported that the BMET is able to differentiate between clients with stroke vs. healthy adults.
– Three studies reported that the VMET is able to differentiate between clients with stroke vs. healthy adults, and between healthy older adults vs. healthy younger adults.

Sensitivity/Specificity:
– One study reported 85% sensitivity and 95% specificity when using a cut-off score ≥ 7 errors on the MET-HV with clients with chronic acquired brain injury including stroke.
– One study reported 82% sensitivity and 95.3% specificity when using a cut-off score ≥ 12 errors on the MET-SV with clients with brain injury including stroke.

Floor/Ceiling Effects No studies have reported on the floor/ceiling effects of the MET.
Does the tool detect change in patients? Responsiveness of the MET has not been formally evaluated, however:
– One study used a modified version of the MET-HV and MET-SV to measure change following intervention;
– One study used the MET-HV and the VMET to detect change in multi-tasking skills of clients with stroke following intervention.
Acceptability The MET provides functional assessment of executive function as it enables clients to participate in real-world activities.
Feasibility Administration of the MET requires access to a shopping area and so is not always feasible in a typical clinical setting. Some tasks may need to be adapted depending on the rehabilitation setting. Administration time can be lengthy. Ecological validity is supported.
How to obtain the tool? Baycrest MET is available from the author: ddawson@research.baycrest.org

Psychometric Properties

Overview

A literature search was conducted to identify publications on the psychometric properties of the Multiple Errands Test (MET) relevant to a population of patients with stroke. Of the 10 studies reviewed, 8 included a mixed population of patients with acquired brain injury including stroke. Studies have reviewed psychometric properties of the original MET, Hospital Version (MET-HV), Simplified Version (MET-SV), Baycrest MET (BMET) and Virtual MET (VMET), as indicated in the summaries below. While research indicates that the MET demonstrates adequate validity and reliability in populations with acquired brain injury including stroke, further research regarding responsiveness of the measure is warranted.

Floor/Ceiling Effects

No studies have reported on floor/ceiling effects of the MET with a stroke population.

Reliability

Internal consistency:
Knight, Alderman & Burgess (2002) calculated internal consistency of the MET-HV in a sample of 20 patients with chronic acquired brain injury (traumatic brain injury, n=12; stroke, n=5, both TBI and stroke, n=3) and 20 healthy control subjects matched for gender, age and IQ, using Cronbach’s alpha. Internal consistency was adequate (α=0.77).

Test-retest:
No studies have reported on the test-retest reliability of the MET.

Intra-rater:
No studies have reported on the intra-rater reliability of the MET.

Inter-rater:
Knight, Alderman & Burgess (2002) calculated inter-rater reliability of the MET-HV error categories in a sample of 20 patients with chronic acquired brain injury (traumatic brain injury, n=12; stroke, n=5, both TBI and stroke, n=3) and 20 healthy control subjects matched for gender, age and IQ, using intraclass correlation coefficients. Participants were scored by 2 assessors. Inter-rater reliability was excellent (ICC ranging from 0.81-1.00). The ‘rule breaks’ error category demonstrated the strongest inter-rater reliability (ICC=1.00).

Dawson, Anderson, Burgess, Cooper, Krpan and Stuss (2009) examined inter-rater reliability of the BMET with clients with stroke (n=14) or traumatic brain injury (n=13) and healthy matched controls (n=25), using Intraclass Correlation Coefficients and 2-way random effects models. Participants were scored by two assessors. Inter-rater reliability was adequate to excellent for the five summary measures used: mean number of tasks completed accurately (ICC = 0.80), mean number of rules adhered to (ICC = 0.71), mean number of total errors (ICC = 0.82), mean number of total rules broken (ICC = 0.88) and mean number of requests for help (ICC = 0.71).

Validity

Content:

Shallice & Burgess (1991) evaluated the MET in a sample of 3 patients with traumatic brain injury (TBI) who demonstrated above-average performance on measures of general ability and normal or near-normal performance on frontal lobe tests, and 9 age- and IQ-matched controls. Participants were monitored by two observers and were scored according to number of errors (inefficiencies, rule breaks, interpretation failures, task failures and total score) and qualitative observation. The patients demonstrated qualitatively and quantitatively impaired performance, particularly relating to rule breaks and inefficiencies. The most difficult subtest was the least sensitive part of the procedure and presented difficulties for both patients and control subjects.

Criterion:

Concurrent:
No studies have reported on the concurrent validity of the MET in a stroke population.

Predictive:
Maier, Krauss & Katz (2011) examined predictive validity of the MET-HV in relation to community participation with a sample of 30 patients with acquired brain injury including stroke (n=19). Community participation was measured using the Mayo-Portland Adaptability Inventory (MPAI-4) Participation Index (M2PI), completed by the participant and a significant other. The MET-HV was administered 1 week prior to discharge from rehabilitation and the M2PI was administered at 3 months post-discharge. Analyses were performed using Pearson correlation analysis and partial correlation controlling for cognitive status using FIM Cognitive scores. Predictably, higher MET-HV error scores correlated with more restrictions in community participation. There were adequate correlations between participants’ and significant others’ M2PI total score and MET-HV total error score (r = 0.403, 0.510 respectively), inefficiencies (r = 0.353, 0.524 respectively) and rule breaks (r = 0.361, 0.449 respectively). The ability for the MET total error score to predict the M2PI significant other score remained significant but poor following partial correction controlling for cognitive status using FIM Cognitive scores (r = 0.212).

Construct:

Convergent/Discriminant:
Knight, Alderman & Burgess (2002)* examined convergent validity of the MET-HV by comparison with tests of IQ and cognitive functioning, traditional frontal lobe tests and ecologically sensitive executive function tests, in a sample of 20 patients with chronic acquired brain injury (traumatic brain injury, n=12; stroke, n=5, both TBI and stroke, n=3). Tests of IQ and cognitive functioning included the National Adult Reading Test – Revised Full Scale Intelligence Quotient (NART-R FSIQ), Weschler Adult Intelligence Scale – Revised Full Scale Intelligence Quotient (WAIS-R FSIQ), Adult Memory and Information Processing Battery (AMIPB), Rivermead Behavioural Memory Test (RBMT) and Visual Objects and Space Perception battery (VOSP). Frontal lobe tests included verbal fluency, the Cognitive Estimates Test (CET), Modified Card Sorting Test (MCST), Tower of London Test (TOLT) and versions of the hand manipulation and hand alternation tests. Ecologically sensitive executive function tests included the Behavioural Assessment of the Dysexecutive Syndrome battery (BADS) and the Test of Everyday Attention (TEA) Map Search and Visual Elevator tasks. The Dysexecutive (DEX) questionnaire was also used, although proxy reports were used rather than self-reports due to identified lack of insight of individuals with brain injury. There were excellent correlations between the MCST percentage perseverative errors with MET-HV rule breaks (r=0.66) and MET-HV total errors (r=0.67) following Bonferroni adjustment. There were excellent correlations between the BADS Profile score and the MET-HV task failures (r = -0.58), interpretation failures (r = 0.64) and total errors (r = -0.57). There was an excellent correlation between the DEX intentionality factor and MET-HV task failures (r = 0.70). In addition, the relationship between the MET-HV and DEX was re-evaluated to control for possible confounding effects; controlling variables age, familiarity and memory function with respect to MET-HV task failures resulted in excellent correlations with the DEX total score (r = 0.79) and DEX inhibition (r = 0.69), intentionality (r = 0.76) and executive memory (r = 0.67) factors. There was an adequate correlation between the RBMT Profile Score and the MET-HV number of task failures (r=-0.57). There were no significant correlations between the MET and other tests of IQ and cognitive functioning (MET-HV, NART-R FSIQ, WAIS-R FSIQ, AMIPB, VOSP), and other frontal lobe tests (verbal fluency, CET, TOLT, hand manipulation and hand alternation tests), other ecologically sensitive executive function tests (TEA Map Search and Visual Elevator tasks) or other DEX factors (positive affect, negative affect).
Note: Initial correlations were measured using Pearson correlation coefficient and significance levels were subsequently adjusted by Bonferroni adjustment to account for multiple comparisons; results reported indicate significant correlations following Bonferroni adjustment.

Rand, Rukan, Weiss & Katz (2009a)* examined convergent validity of the MET-HV by comparison with measures of executive function and IADLs with a sample of 9 patients with subacute or chronic stroke, using Spearman correlation coefficients. Executive function was measured using the BADS Zoo Map test and IADLs were measured using the IADL questionnaire. There were excellent negative correlations between the BADS Zoo Map and MET-HV outcome measures of total number of mistakes (r = -0.93), partial mistakes in completing tasks (r = -0.80), non-efficiency mistakes (r = -0.86) and time to complete the MET (r = -0.79). There were excellent correlations between the IADL questionnaire and the MET-HV number of mistakes of rule breaks (r = 0.80) and total number of mistakes (r = -0.76).

Maier, Krauss & Katz (2011)* examined convergent validity of the MET-HV by comparison with the FIM Cognitive score with a sample of 30 patients with acquired brain injury including stroke (n=19), using Pearson correlation analysis. There was an excellent negative correlation between MET-HV total errors score and FIM Cognitive score (r = -0.67).

Alderman, Burgess, Knight and Henman (2003)* examined convergent validity of the MET-SV by comparison with tests of IQ, executive function and everyday executive abilities with 50 clients with brain injury including stroke (n=9). Neuropsychological tests included the WAIS-R FSIQ, BADS, Cognitive Estimates Test, FAS verbal fluency test, a modified version of the Wisconsin Card Sorting Test (MWCST) and the DEX. There were adequate correlations between MET-SV task failure errors and WAIS-R FSIQ (r = -0.32), MWCST perseverative errors (r = 0.39), BADS profile score (r = -0.46) and Zoo-Map (r = -0.46) and Six Element Test (r = -0.41) subtests. There were adequate negative correlations between MET-SV social rule breaks and the Cognitive Estimates (r = -0.33), and between MET-SV task rule breaks, social rule breaks and total rule breaks and the BADS Action Program subtest (r = -0.42, -0.40, -0.43 respectively). There were poor to adequate negative correlations between the DEX and MET-SV rule breaks (r = -0.30), task failures (r = -0.25) and total errors (r = -0.37).

In a subgroup analysis of individuals with brain injury who passed traditional executive function tests but failed the MET-SV (n=17), there were adequate to excellent correlations between MET-SV inefficiencies and DEX factors of intentionality and negative affect (r = 0.59, -0.76); MET-SV interpretation failures and DEX inhibition and total (r = -0.67, -0.57); MET-SV total and actual rule breaks and DEX inhibition (r = -0.70, 0.66), intentionality (r = 0.60, 0.64) and total (r = -0.57, 0.59); MET-SV social rule breaks and DEX positive and negative affect (r = 0.79, -0.59); MET-SV task failures and DEX inhibition and positive affect (r = -0.58, -0.52), and MET-SV total errors and DEX intentionality (r = 0.67).

Dawson et al. (2009)* examined convergent validity of the BMET by comparison with other measures of IADL and everyday function with 14 clients with stroke, using Pearson correlation. Other measures included the DEX (significant other report), Stroke Impact Profile (SIP), Assessment of Motor and Process Skills (AMPS) and Mayo Portland Adaptability Inventory (MPAI) (significant other report). There were excellent correlations between the BMET number of rules broken and the SIP – Physical (r = 0.78) and Affective behavior (r = 0.64) scores and the AMPS motor score (r = -0.75). There was an adequate correlation between the BMET time to completion and SIP physical score (r = 0.54).

Rand et al. (2009a)* examined convergent validity of the VMET by comparison with the BADS Zoo Map test and IADL questionnaire with the same sample of 9 patients with subacute or chronic stroke, using Spearman correlation coefficients. There was an excellent negative correlation between the BADS Zoo Map and VMET outcome measure of non-efficiency mistakes (r = -0.87), and between the IADL and VMET total number of mistakes (r = -0.82).

Rand et al. (2009a) also examined the relationships between the scores of the VMET and those of the MET-HV using Spearman and Pearson correlation coefficients. Among patients with stroke, there were excellent correlations between MET-HV and VMET outcomes for the total number of mistakes (r = 0.70), partial mistakes in completing tasks (r = 0.88) and non-efficiency mistakes (r = 0.73). Analysis of the whole population indicated adequate to excellent correlations between MET-HV and VMET outcomes for the total number of mistakes (r = 0.77), complete mistakes of completing a task (r = 0.63), partial mistakes in completing tasks (r = 0.80), non-efficiency mistakes (r = 0.72) and use of strategies (r = 0.44), but not for rule break mistakes.

Raspelli et al. (2010) examined convergent validity of the VMET by comparison with neuropsychological tests, with 6 clients with stroke and 14 healthy subjects. VMET outcome measures included time, searched item in the correct area, sustained attention, maintained sequence and no perseveration. Neuropsychological tests included the Trail Making Test, Corsi spatial memory supra-span test, Street’s Completion Test, Semantic Fluencies and Tower of London test. There were excellent correlations between the VMET variable ‘time’ and the Semantic Fluencies test (r = -0.87) and the Tower of London test (r = -0.82); between the VMET variable ‘searched item in the correct area’ and the Trail Making Test (r = 0.96); and between the VMET variables ‘sustained attention’, ‘maintained sequence’ and ‘no perseveration’ and Corsi’s supra-span test (r = 0.84) and Street’s Completion Test (r = -0.86).

Raspelli et al. (2012) examined convergent validity of the VMET by comparison with the Test of Attentional Performance (TEA) with 9 clients with stroke. VMET outcome measures included time, errors, inefficiencies, rule breaks, strategies, interpretation failures and partial-task failures. Authors reported excellent correlations between the VMET outcomes time, inefficiencies and total errors and TEA tests (range r = -0.67 to 0.81).
Note: Other neuropsychological tests were administered but correlations are not reported (Mini Mental Status Examination (MMSE), Beck Depression Inventory (BDI), State and Trait Anxiety Index (STAI), Behavioural Inattention Test (BIT) – Star Cancellation Test, Brief Neuropsychological Examination (ENB) – Token Test, Street’s Completion Test, Stroop Colour-Word Test, Iowa Gambling Task, DEX and ADL/IADL Tests).
*Note: The correlations between the MET and other measures of everyday executive functioning and IADLs also provide support for the ecological validity of the MET (as reported by the authors of these articles).

Known Group:
Knight, Alderman & Burgess (2002) examined known-group validity of the MET-HV in a sample of 20 patients with chronic acquired brain injury (traumatic brain injury, n=12; stroke, n=5, both TBI and stroke, n=3) and 20 healthy control subjects (hospital staff members) matched for gender, age and IQ*. Clients with brain injury made significantly more rule breaks (p=0.002) and total errors (p<0.001), and achieved significantly fewer tasks (p<0.001) than control subjects. Clients with brain injury used significantly more strategies such as looking at a map (p=0.008), reading signs (p=0.006), although use of strategies had little effect on test performance. The test was able to discriminate between individuals with acquired brain injury and healthy controls.
*Note: IQ was measured using the National Adult Reading Test – Revised Full Scale Intelligence Quotient (NART-R FSIQ).

Rand et al. (2009a) examined known group validity of the MET-HV with 9 patients with subacute or chronic stroke, 20 healthy young adults and 20 healthy older adults, using Kruskal Wallis H. Patients with stroke made more mistakes than older adults on VMET outcomes of total mistakes, mistakes in completing tasks, partial mistakes in completing tasks and non-efficiency mistakes, but not rule break mistakes or use of strategies mistakes. Older adults made more mistakes than younger adults on VMET outcomes of total mistakes, partial mistakes in completing tasks and non-efficiency mistakes, but not mistakes in completing tasks, rule break mistakes or use of strategies mistakes.

Alderman et al. (2003) examined known group validity of the MET-SV with 46 individuals with no history of neurological disease (hospital staff members) and 50 clients with brain injury including stroke (n=9), using a series of t-tests. Clients with brain injury made significantly more rule breaks (t = 4.03), task failures (t = 10.10), total errors (t = 7.18), and social rule breaks (chi square 4.3) than individuals with no history of neurological disease. Results regarding errors were preserved when group comparisons were repeated using age, familiarity and cognitive ability (measured by the NART-R FSIQ) as covariates (F = 11.79, 40.82, 27.92 respectively). There was a significant difference in task failures between groups after covarying for age, IQ (measured by the WAIS-R FSIQ) and familiarity with the shopping centre (F = 11.57). Clients with brain injury made approximately three times more errors as healthy individuals. For both groups, rule breaks and task failures were the most common errors.

Dawson et al. (2009) examined known group validity of the BMET with 14 clients with stroke and 13 healthy matched controls, using a series of t-tests. Clients with stroke performed significantly worse on number of tasks completed accurately (d = 0.84, p<0.05), rule breaks (d = 0.92, p<0.05) and total failures (d = 1.05, r<0.01). The proportion of group members who completed fewer than 40% (< 5) tasks satisfactorily was also significantly different between the two groups (28% of clients with stroke vs. 0% of healthy matched controls, p<0.05).
Note: d is the effect size; effect sizes ≥ 0.7 are considered large.

Rand et al. (2009a) examined known group validity of the VMET with a sample of 9 patients with subacute or chronic stroke, 20 healthy young adults and 20 healthy older adults, using Kruskal Wallis H. Patients with stroke made more mistakes than older adults on all VMET outcomes except for rule break mistakes. Older adults made more mistakes than young adults on all VMET outcomes except for the use of strategies mistakes.

Raspelli et al. (2010) examined known group validity of the VMET with 6 clients with stroke and 14 healthy subjects. There were significant differences between groups in time taken to execute the task (higher for healthy subjects) and in the partial error ‘Maintained task objective to completion’.

Raspelli et al. (2012) examined known group validity of the VMET with 9 clients with stroke, 10 healthy young adults and 10 healthy older adults, using Kruskal-Wallis procedures. Results showed that clients with stroke scored lower in VMET time and errors than older adults, and that older adults scored lower in VMET time and errors than young adults.

Sensitivity/ Specificity:
Knight, Alderman & Burgess (2002) investigated sensitivity and specificity of the MET-HV in a sample of 20 patients with chronic acquired brain injury (traumatic brain injury, n=12; stroke, n=5, both TBI and stroke, n=3) and 20 healthy control subjects matched for gender, age and IQ*. A cut-off score ≥ 7 errors (i.e. 5th percentile of total errors of control subjects) resulted in correct identification of 85% of participants with acquired brain injury (85% sensitivity, 95% specificity).
*Note: IQ was measured using the National Adult Reading Test – Revised Full Scale Intelligence Quotient (NART-R FSIQ).

Alderman et al. (2003) reported on sensitivity and specificity of the MET-SV with 46 individuals with no history of neurological disease and 50 clients with brain injury including stroke (n=9). Using a cutoff score ≥ 12 errors (i.e. 5th percentile of controls) results in 44% sensitivity (i.e. correct classification of clients with brain injury) and 95.3% specificity (i.e. correct classification of healthy individuals). The authors caution that deriving a single measure based only on number of errors fails to consider between-group qualitative differences in performance. Accordingly, error scores were recalculated to reflect “normality” of the error type, with weighting of errors according to prevalence in the healthy control group (acceptable errors seen in up to 95% of healthy controls = 1; errors demonstrated by ≥ 5% of healthy controls = 2; errors unique to the patient group = 3). Using a cutoff score ≥ 12 errors (5th percentile of controls) resulted in 82% sensitivity and 95.3% specificity. The MET-SV was more sensitive than traditional tests of executive function (Cognitive Estimates, FAS Verbal Fluency, MWCST), and MET-SV error category scores were highly predictive of rating s of executive symptoms of patients who passed traditional executive function tests but failed the MET-SV shopping task.

Responsiveness

Two studies used the MET (MET-HV, VMET and modified version of the MET-HV & MET-SV) to measure change following intervention.

Novakovic-Agopian et al. (2011) developed a modified version of the MET-HV and MET-SV to be used in local hospital settings. They developed 3 alternate forms that were used in a pilot study examining the effect of goal-oriented attentional self-regulation training with a sample of 16 patients with chronic brain injury including stroke or cerebral hemorrhage (n=3). A pseudo-random crossover design was used. During the first 5 weeks, one group (Group A) completed goal-oriented attentional self-regulation training while the other group (Group B) only received a 2-hour educational instructional session. In the subsequent phase, conditions were switched such that participants in Group B received goals training for 5 weeks while those in Group A received educational instruction. At week 5 the group that received goal training first demonstrated a significant reduction in task failures (p<0.01), whereas the group that received the educational session demonstrated no significant improvement in MET scores. From week 5 to week 10 there were no significant changes in MET scores in either group.

Rand, Weiss and Katz (2009b) used the MET-HV and VMET to detect change in multi-tasking skills of 4 clients with subacute stroke following virtual reality intervention using the VMall virtual supermarket. Clients demonstrated improved performance on both measures following 3 weeks of multi-tasking training using the VMall virtual supermarket.

References

  • Alderman, N., Burgess, P.W., Knight, C., & Henman, C. (2003). Ecological validity of a simplified version of the multiple errands shopping test. Journal of the International Neuropsychological Society, 9, 31-44.
  • Dawson, D.R., Anderson, N.D., Burgess, P., Cooper, E., Krpan, K.M., & Stuss, D.T. (2009). Further development of the Multiple Errands Test: Standardized scoring, reliability, and ecological validity for the Baycrest version. Archives of Physical Medicine and Rehabilitation, 90, S41-51.
  • Knight, C., Alderman, N., & Burgess, P.W. (2002). Development of a simplified version of the Multiple Errands Test for use in hospital settings. Neuropsychological Rehabilitation, 12(3), 231-255.
  • Maier, A., Krauss, S., & Katz, N. (2011). Ecological validity of the Multiple Errands Test (MET) on discharge from neurorehabilitation hospital. Occupational Therapy Journal of Research: Occupation, Participation and Health, 31(1) S38-46.
  • Novakovic-Agopian, T., Chen, A.J.W., Rome, S., Abrams, G., Castelli, H., Rossi, A., McKim, R., Hills, N., & D’Esposito, M. (2011). Rehabilitation of executive functioning with training in attention regulation applied to individually defined goals: A pilot study bridging theory, assessment, and treatment. The Journal of Health Trauma Rehabilitation, 26(5), 325-338.
  • Novakovic-Agopian, T., Chen, A. J., Rome, S., Rossi, A., Abrams, G., Dʼesposito, M., Turner, G., McKim, R., Muir, J., Hills, N., Kennedy, C., Garfinkle, J., Murphy, M., Binder, D., Castelli, H. (2012). Assessment of Subcomponents of Executive Functioning in Ecologically Valid Settings: The Goal Processing Scale. The Journal of Health Trauma Rehabilitation, 2012 Oct 16. [Epub ahead of print]
  • Rand, D., Rukan, S., Weiss, P.L., & Katz, N. (2009a). Validation of the Virtual MET as an assessment tool for executive functions. Neuropsychological Rehabilitation, 19(4), 583-602.
  • Rand, D., Weiss, P., & Katz, N. (2009b). Training multitasking in a virtual supermarket: A novel intervention after stroke. American Journal of Occupational Therapy, 63, 535-542.
  • Raspelli, S., Carelli, L., Morganti, F., Poletti, B., Corra, B., Silani, V., & Riva, G. (2010). Implementation of the Multiple Errands Test in a NeuroVR-supermarket: A possible approach. Studies in Health Technology and Informatic, 154, 115-119.
  • Raspelli, S., Pallavicini, F., Carelli, L., Morganti, F., Pedroli, E., Cipresso, P., Poletti, B., Corra, B., Sangalli, D., Silani, V., & Riva, G. (2012). Validating the Neuro VR-based virtual version of the Multiple Errands Test: Preliminary results. Presence, 21(1), 31-42.
  • Shallice, T. & Burgess, P.W. (1991). Deficits in strategy application following frontal lobe damage in man. Brain, 114, 727-741.

See the measure

How to obtain the Multiple Errands Test?

See the papers below for test instructions of the Simplified Version (MET-SV) and the Hospital Version (MET-HV):

  • Alderman, N., Burgess, P.W., Knight, C., & Henman, C. (2003). Ecological validity of a simplified version of the multiple errands shopping test.Journal of the International Neuropsychological Society, 9, 31-44.
  • Knight, C., Alderman, N., & Burgess, P.W. (2002). Development of a simplified version of the Multiple Errands Test for use in hospital settings.Neuropsychological Rehabilitation, 12(3), 231-255.

The Baycrest MET can be obtained by contacting the author: ddawson@research.baycrest.org

Table of contents

Reintegration to Normal Living Index (RNLI)

Evidence Reviewed as of before: 19-08-2008
Author(s)*: Elissa Sitcoff, BA BSc
Editor(s): Nicol Korner-Bitensky, PhD OT; Lisa Zeltzer, MSc OT

Purpose

The Reintegration to Normal Living Index (RNLI) was developed to assess, quantitatively, the degree to which individuals who have experienced traumatic or incapacitating illness achieve reintegration into normal social activities (e.g. recreation, movement in the community, and interaction in family or other relationships). Reintegration to normal living was defined by the scale authors as the “reorganization of physical, psychological, and social characteristics of an individual into a harmonious whole so that one can resume well-adjusted living after incapacitating illness or trauma” (Wood-Dauphinee & Williams, 1987).

The RNLI has been tested for use with individuals with stroke, malignant tumors, degenerative heart disease, central nervous system disorders, arthritis, fractures and amputations; spinal cord injury; traumatic brain injury; rheumatoid arthritis; subarachnoid hemorrhage; hip fracture; physical disability; and community-dwelling elderly.

In-Depth Review

Purpose of the measure

The Reintegration to Normal Living Index (RNLI) was developed to assess, quantitatively, the degree to which individuals who have experienced traumatic or incapacitating illness achieve reintegration into normal social activities (e.g. recreation, movement in the community, and interaction in family or other relationships). Reintegration to normal living was defined by the scale authors as the “reorganization of physical, psychological, and social characteristics of an individual into a harmonious whole so that one can resume well-adjusted living after incapacitating illness or trauma” (Wood-Dauphinee & Williams, 1987).

The RNLI has been tested for use with individuals with stroke, malignant tumors, degenerative heart disease, central nervous system disorders, arthritis, fractures and amputations; spinal cord injury; traumatic brain injury; rheumatoid arthritis; subarachnoid hemorrhage; hip fracture; physical disability; and community-dwelling elderly.

Available versions

The RNLI was developed by Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer in 1988.

Features of the measure

Items:

The RNLI index is made up of 11 declarative statements (e.g. I move around my living quarters as I feel necessary), including the following domains: indoor, community, and distance mobility; self-care; daily activity (work and school); recreational and social activities;; family role(s); personal relationships; presentation of self to others and general coping skills. The first 8 items represent ‘daily functioning’ and the remaining 3 items represent ‘perception of self’.

Scoring:

Each domain is accompanied by a visual analogue scale (VAS) (0 to 10 cm). The VAS is anchored by the statements “does not describe my situation” (1 or minimal integration) and “fully describes my situation” (10 or complete integration). Individual item scores are summed to provide a total score out of 110 points that is proportionally converted to create a score out of 100.

Three- and 4-point categorical scoring systems were also developed (Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer, 1988), and the 3-point categorical system has been used in the evaluation of stroke patients (Mayo et al., 2000; Mayo et al., 2002). In the 3-point system, an additional category is inserted between the two anchor points (“partially describes my situation”) and the respondent selects the most applicable of the three categories. This option yields total scale scores from 22-0, with higher scores indicating poorer reintegration (Mayo et al., 2000, Mayo et al., 2002).

Time:

The time to administer depends on the mode of administration (e.g. self-administration, interviewer-administration, proxy, postal, etc.) and the participant’s abilities, but typically takes less than 10 minutes to complete.

Subscales:

There are two subscales to the RNLI: Daily Functioning (indoor, community, and distance mobility; self-care; daily activity (work and school); recreational and social activities; general coping skills) and Perception of Self (family role(s); personal relationships; and presentation of self to others.).

Equipment:

Only the test and a pencil are required to complete the RNLI.

Training:

The RNLI requires no training to administer.

Alternative forms of the Reintegration to Normal Living Index

  • Reintegration to Normal Living Index – Postal Version (RNLI-P) was developed by Daneski, Coshall, Tilling, and Wolfe in 2003.
    This measure modified the original RNLI in phrasing and scoring for use by post with stroke patients. The RNLI – P uses an agree/disagree format (0=disagree, 1= agree).
  • There are also versions created with minor modifications in wording to the original RNLI for: individuals who use adaptive devices motor aids or human assistance where the use of equipment and resources are clarified; use by health care professionals; and use by significant others (Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer,1988).

Client suitability

Can be used with:

  • Patients with stroke.

Should not be used with:

  • The use of a visual analogue scale may not be appropriate for the assessment of some stroke patients (i.e. those with attentional deficits or visual impairments or difficulty comprehending the meaning of a VAS). Instead, the use of the 3- or 4-point categorical scoring system is recommended.

In what languages is the measure available?

The RNLI is available in Canadian English and Canadian French (Wood-Dauphinee & Williams, 1987; Wood-Dauphinee, Opzoomer, Williams, Marchand, & Spitzer, 1988).

Please click here to access the french language version. A research version is also available.

Summary

What does the tool measure? The degree to which individuals who have experienced traumatic or incapacitating illness achieve reintegration into normal social activities.
What types of clients can the tool be used for? The RNLI has been tested for use with individuals with stroke, malignant tumors, degenerative heart disease, central nervous system disorders, arthritis, fractures and amputations; spinal cord injury; traumatic brain injury; rheumatoid arthritis; subarachnoid hemorrhage; hip fracture; physical disability; and community-dwelling elderly.
Is this a screening or assessment tool? Assessment
Time to administer The amount of time it takes to administer the RNLI is dependent upon mode of administration and participant’s abilities but should take approximately 10 minutes.
Versions
  • Reintegration to Normal Living Index ( RNLI)
  • Reintegration to Normal Living Index- Postal Version (RNLI-P).
  • There are also versions created with minor modifications in wording to the original RNLI for: individuals who use adaptive devices motor aids or human assistance where the use of equipment and resources are clarified; use by health care professionals; and use by significant others.
  • The original RNLI index is made up of 11 declarative statements Three- and 4-point categorical scoring systems are also available.
Other Languages Canadian French (Please click here to access the french language version. A research version is also available.)
Measurement Properties
Reliability Internal consistency:
Six studies have examined the internal consistency of the RNLI. Four reported excellent Cronbach’s alphas. One reported excellent Cronbach alphas for the total RNLI patient and significant other score as well as for the patient score on the Perception of Self subscale, adequate Cronbach alphas for the Daily Functioning subscale for both patient and significant other score, as well as on the significant other score on the Perception of Self subscale. One study reported adequate to excellent Cronbach alphas.

Test-retest:
Three studies have examined the test-retest reliability of the RNLI and reported adequate test-retest agreement between items using kappa statistics, and excellent test-retest on the global score using correlation coefficients.

Intra-rater:
No studies have examined the intra-rater reliability of the RNLI.

Inter-rater:
No studies have examined the inter-rater reliability of the RNLI.

Validity Construct:
Convergent/Discriminant:
– Excellent correlations between the total score of the RNLI-P and the Frenchay Activities Index (FAI), the Short Form 36 Health Survey (SF-36) and with the Hospital Anxiety and Depression Scale-Depression subscale (HADS). Excellent correlations between the Daily Function subscale of the RNLI-P and the FAI and the SF-36. Poor correlations between the RNLI-P Daily Functioning subscale and the HADS-Anxiety subscale as well as between the Perceptions of Self subscale and both the FAI and the Barthel Index.
– Excellent correlation between the RNLI and the Quality of Life Index (QL) and with a measure of psychological wellbeing. Excellent correlation between Daily Functioning subscale and with Quality of Life Index items Activity and Daily Living. Adequate correlations between Perceptions of Self subscale and Support and Outlook items from the Quality of Life Index. Strong correlation between the RNLI and the Participation Survey/Mobility (PARTS/M). A positive relationship between the Health Options Scale and the RNLI for stroke survivors well as a positive relationship between the Herth Hope Index and the RNLI for both stroke survivors and their spouses.
– Adequate correlation between the RNLI and items on the subscale related to physical performance of the Prosthetic Profile of the Amputee (PPA) with the exception of the item “active use of the prosthesis indoors” which was poor. No correlation between items of the Perception of Self subscale of the RNLI with items on the subscale related to physical performance of the PPA with the exception of prosthetic wear which was adequate. Adequate to excellent correlations between items of the total RNLI with items in the subscale related to Physical performance of the PPA with the exception of items “Active use indoors” and “Active use outdoors” which had non-significant correlations.
– Poor to adequate correlations between items of the total RNLI, and its two subscales with items on the PPA subscale related to acceptance of amputation and prosthesis.
– Significant correlations between the RNLI and the Functional Independence Measure (FIM).
– Adequate to excellent correlations between scores the total RNLI and the Daily Functioning subscale with patient (with Rheumatoid arthritis) age, number of affected joints, the Functional Independence Measure (FIM), the Lee Index (pain, fatigue, and stiffness), and the American Rheumatism Association Classification. The total RNLI was also adequately correlated to disease duration.
Acceptability The use of the 3 or 4 point categorical scoring system may be more appropriate for the assessment of some stroke patients than the visual analogue scale
Feasibility The administration of the RNLI is quick and simple and requires no training to administer. The RNLI index is made up of 11 declarative statements representing the domains ‘daily functioning’ (indoor, community, and distance mobility; self-care; daily activity (work and school); recreational and social activities;; family role(s); personal relationships; and ‘perception of self'(presentation of self to others, general coping skills. Each domain is accompanied by a visual analogue scale (VAS) (0 to 10 cm). The VAS is anchored by the statements “does not describe my situation” (1 or minimal integration) and “fully describes my situation” (10 or complete integration). Individual item scores are summed to provide a total score out of 110 points that is proportionally converted to create a score out of 100.
How to obtain the tool? The RNLI is available by clicking here.

Psychometric Properties

Overview

We conducted a literature search to identify all relevant publications on the psychometric properties of the Reintegration to Normal Living Index (RNLI).

Floor/Ceiling Effects

Not yet examined.

Reliability

Internal consistency:
Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer (1988) administered the RNLI to three samples of patients with varied diagnoses to determine internal consistency. The RNLI was completed by patients, significant others, and healthcare professionals. The Cronbach’s alphas were excellent for patients, significant others, and health care professionals (alpha = 0.90, 0.92, and 0.95, respectively). Corrected item to total correlations ranged from 0.39 (patient assessment of “comfort with self-care needs”) to 0.75 for patients, 0.61 to 0.87 for significant others, and 0.70 to 0.90 for health professionals.

Tooth, McKenna, Smith, and O’Rourke (2003) administered the RNLI to 57 pairs of patients and significant others six months after stroke rehabilitation. Cronbach’s alphas were excellent for the total RNLI patient and significant other scores (alpha = 0.80 and 0.81, respectively). For the Daily Functioning subscale, adequate Cronbach’s alphas were found for both patient and significant other scores (alpha = 0.71 and 0.73, respectively). For the Perception of Self subscale, Cronbach’s alpha was excellent for patient scores (alpha = 0.84) and adequate for significant other scores (alpha = 0.76).

Steiner et al. (1996) examined the internal consistency of the RNLI in two samples of community-dwelling persons aged 75 and over (n=414, n=50). Cronbach’s alphas were adequate (0.76) to excellent (0.83).

Daneski, Coshall, Tilling, and Wolfe (2002) examined the internal consistency of a postal version of the RNLI (the RNLI-P) administered to 76 patients with stroke (at one-year). The Cronbach’s alpha was excellent (0.84).

Stark, Edwards, Hollingsworth, and Gray (2005) administered the RNLI to 604 people between the ages of 18 and 80 years who had a mobility limitation (including patients with spinal cord injury, Multiple Sclerosis, stroke, cerebral palsy, and polio), lived in the community, and had been discharged from rehabilitation for at least 1 year. The Cronbach’s alpha for this sample was excellent (0.91).

Bluvol and Ford-Gilboe (2004) administered the RNLI to both spouses in 40 families in which one of the partners had experienced a stroke with moderate to severe functional impairments (6 months to 5 years post-stroke). The internal consistency of the measure was excellent for both the partners with stroke (alpha = 0.92) and their spouses (alpha = 0.85).

Test-retest:
Steiner et al. (1996) examined the test-retest reliability of the RNLI in 50 community-dwelling persons aged 75 and over interviewed twice, by the same interviewer, with 7 to 14 days between interviews. Test-retest for the total sample of community-dwelling elderly was excellent (r = 0.83). When examined by age group, correlations were excellent for the 75 to 79 age group (r = 0.82), 80 to 84 age group (r = 0.93), and for the 85+ age group (r = 0.76).

Daneski, Coshall, Tilling and Wolfe (2002) examined the test-retest reliability of a postal version of the RNLI (the RNLI-P) in 26 patients with stroke (3-12 months post-stroke) who completed the test twice within a 2-week interval. All 11 items demonstrated agreement between the two occasions above that expected by chance. Kappa values ranged from poor to excellent agreement (kappa = 0.38 for the item “embarrassed when with others”, to 0.92 for the item “getting around outside”).

Korner-Bitensky, Wood-Dauphinee, Siemiatycki, Shapiro, and Becker (1994) examined the test-retest reliability of the RNLI in 366 patients with a diagnosis of stroke or orthopedic condition discharged from a rehabilitation hospital. The test was administered twice – once by face-to-face interview and once by a structured telephone interview to either a self or proxy respondent. The interclass coefficient (ICC) for the RNL Index was 0.80 indicating excellent agreement between the two modes of interview. However, for the self-respondents, poor community reintegration was reported more often during the home interview than the interview conducted over the telephone.

Type of rater:
Korner – Bitensky, Wood Dauphinee, Shapiro, and Becker (1994) analyzed the reliability of RNLI scores of 366 participants (with stroke or an orthopedic condition post discharge from a rehabilitation hospital) who completed both a home interview (conducted by a health professional only) and a telephone interview (conducted by either a lay person or health professional). Results revealed that there were no significant differences on the comparison of kappa scores when patients were interviewed by lay interviewers or health professionals. When a dichotomized score of 40 was used (0-40 = no disability, scores of >40 equals disability), the group interviewed by phone by a layperson was significantly more likely to report difficulties in community reintegration compared to when interviewed face-to-face.

Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer (1988) analyzed the reliability of RNLI scores between patients and relatives and between patients and health professionals. Using Pearson’s correlation coefficient to measure reliability they reported adequate significant other to patient correlations of r = 0.62 and r = 0.65 in two different patient/significant other samples. They also reported poor to adequate health professional to patient correlations of r = 0.39 and r = 0.43. Based on these results, the authors stated that patients or significant others could complete the RNLI but that the use of health professionals as proxies should be avoided.

Trombly, Radomski, and Davis (1998) administered the RNLI to 16 adults with traumatic brain injury and their significant others. At admission to a treatment program, patients’ and proxies’ scores did not differ significantly, however at discharge and follow-up, they differed significantly

Tooth, McKenna, Smith, and O’Rourke (2003) examined patient proxy reliability of RNLI scores in 57 subacute patients paired with a significant other 6 months post stroke rehabilitation. Intra-class Correlation Coefficients were poor for the total RNLI score (0.36) and the Daily Functioning subscale (0.24). Adequate reliability was found for the Perception of Self subscale (0.55.).

Validity

Content:

The RNLI was developed based on literature reviews, incorporation of experiences of investigators, and open- and closed-ended questionnaires given to patients with myocardial infarction, cancer, and other chronic diseases, health professionals (physicians, social workers, physical and occupational therapists, psychologists), significant others of patients; and clergy and other lay people.

Construct:

Convergent/Discriminant:
Daneski, Coshall, Tilling and Wolfe (2003) examined the construct validity of a postal version of the RNLI (RNLI-P) with other similar measures in 76 patients with stroke. Excellent correlations were found between the total score on the RNLI-P and the Frenchay Activities Index (FAI – Holbrook & Skilbeck, 1983) (r = 0.69), the Short Form 36 Health Survey (SF-36 – Ware, Snow, Kosinski & Gandek, 1993) (r = 0.74), and with the Hospital Anxiety and Depression Scale-Depression subscale (HADS – Zigmond & Snaith, 1983) (r = -0.61). Excellent correlations were reported between the Daily Function subscale of the RNLI-P and the FAI (r = 0.74) and the SF-36 (r = 0.73). The RNLI-P Daily Function subscale correlated poorly with the HADS-Anxiety subscale(r=-0.30). The Perceptions of Self subscale correlated poorly with the FAI (r = 0.26) and with the Barthel Index (Mahoney & Barthel, 1965), (r = 0.06).

Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer (1988) administered the RNLI to 70 patients with myocardial infarct or cancer and reported excellent correlation with scores on the Quality of Life (QL) Index (Spitzer, Dobson, Hall, Chesterman, Levi, Shepherd, Battista & Catchlove, 1981) (r = 0.68) and with a measure of psychological well-being (r = 0.32 for positive wellbeing, -0.41 for negative wellbeing, and 0.41 for overall). Daily Functioning subscale scores showed excellent correlation with QL Index items Activity and Daily Living (r = 0.67) while Perceptions of Self scores correlated adequately with Support and Outlook from the QL Index (r = 0.36).Items on the QL Index that reflected dimensions not included on the RNLI, correlated less strongly (r < 0.20).

In a study describing the development and psychometric properties of the Participation Survey/Mobility (PARTS/M), Gray, Hollingsworth, Stark and Morgan (2006) administered the RNLI to 604 people with mobility limitations due to a diagnosis of spinal cord injury, Multiple Sclerosis, cerebral palsy, stroke or post poliomyelitis and reported a strong correlation between the two indices (canonical correlation =0.71).

Bluvol and Ford-Gilboe (2004) administered the Herth Hope Index (measure of hope – Herth, 1992), the Health Options Scale (measure of health work – Ford-Gilboe, 1997, 2002b) and the RNLI to both spouses in 40 families in which one of the partners had experienced a stroke with moderate to severe functional impairments (6 months to 5 years post-stroke). They found a positive relationship between the Health Options Scale (health work) and the RNLI for stroke survivors (r = 0.50) but not for their spouses(r = 0.06) as well as a positive relationship between the Herth Hope Index (hope) and the RNLI (quality of life) for both stroke survivors (r = 0.59) and spouses (r = 0.32).
Note: Health work is defined as “an active process through which families learn ways of coping and developing that are conducive to healthy living over time” Ford-Gilboa 2002a.

Gauthier-Gagnon, and Grise (1994) administered the RNLI and the Prosthetic Profile of the Amputee (PPA) questionnaire (Grise, Gauthier-Gagnon, 1993) to 89 people with a lower limb amputation. Items on the Daily Activities subscale of the RNLI correlated adequately (r = 0.36 to 0.56) with items on the subscale related to physical performance of the PPA with the exception of the item “active use of the prosthesis indoors” which was poor ( r = 0.28).

In this same study, items of the Perception of Self subscale of the RNLI failed to correlate with items on the subscale related to physical performance of the PPA with the exception of prosthetic wear which was adequate (r = 0.32).

Items of the total RNLI had adequate to excellent correlations (r = 0.36 to 0.53) with items in the subscale related to Physical performance of the PPA with the exception of items “Active use indoors” and “Active use outdoors” which had non-significant correlations.

Items of the total RNLI, and its two subscales revealed poor to adequate correlations (r = 0.53 to 0.30) with items on the subscale related to acceptance of amputation and prosthesis.

Daverat, Petit, Kemoun, Dartigues, and Barat (1995) conducted a longitudinal study of 149 individuals with long-standing spinal cord injury. The univariate analysis showed that the RNLI significantly correlated with the Functional Independence Measure (FIM) (Hamilton, Granger, & Sherwin, 1987) FIM. The multivariate analysis determined that the following significant seven independent variables contributed to 72% of the RNLI variance. They included the FIM, the Yale Scale Score (Chehrazi, Wagner, Collins, Freedman, 1981) the Centre for Epidemiological Studies Depression Scale (CES-D – Radloff, 1977), living conditions, relationship, sexual life and age.

Calmels, Pereira, Domenach, Pallot-Prades, Alexandre, and Minaire (1994) administered the RNLI to 57 individuals with rheumatoid arthritis, with a mean disease duration of 15 years. In this study, scores on the total RNLI and the Daily Function subscale had adequate to excellent correlations with patient age, number of affected joints, the Functional Independence Measure (FIM), the Lee Index (pain, fatigue, and stiffness), and the American Rheumatism Association Classification (r = 0.38 to 0.84). The total RNLI was also adequately correlated to disease duration (r = 0.31).

McColl, Paterson, Davies, Doubt, and Law (2000) administered the RNLI to 61 community-dwelling individuals with a disability and found that RNLI scores were adequately correlated with the satisfaction subscale of the Canadian Occupational Performance Measure (COPM – Law et al., 1991, 1994, 1998) (r = 0.38) but only poorly correlated with the Performance subscale (r = 0.22). The RNLI had excellent correlations with the Life Satisfaction Scale (Michalos, 1980) (r = 0.71) and with the Satisfaction with Performance Scaled Questionnaire (Yerxa, Burnett-Beaulieu, Stocking & Azen, 1988) (r = 0.72).

Steiner et al. (1996) evaluated the performance of the RNLI in an elderly community-based population (n = 414). The RNLI demonstrated adequate positive correlations with instrumental activities of daily living scale (Lawton, Moss, Fulcomer & Klegan, 1982) (r = 0.47) and perceived health (r = 0.45). Poor to adequate negative correlations were reported for living alone (r = -0.14) and number of both bed days (r = -0.16) and chronic conditions(r = -0.32). There was an unpredicted negative correlation between age and RNLI (r = -0.11).

May and Warren (2002) examined the external and structural components of validity of the spinal cord injury version of the Ferrans and Powers Quality of Life Index (Ferrans & Powers, 1992) in a sample of 98 individuals with spinal cord injury living in the community and reported an excellent correlation with the RNLI (r = -0.65).

Patrick, Perugini, and Leclerc (2002) reported that in a study of 48 consecutive referrals for neuropsychological evaluation following admission to a geriatric rehabilitation inpatient service (for various diagnosis including: orthopedic injury, stroke, functional deconditioning, Parkinson’s disease and other various medical conditions) that the RNLI was significantly correlated to the number of falls sustained and functional status at 6 months. Results of the partial correlations coefficients revealed significant relationships between the RNL and the California Verbal Learning Test (CVLT) (measures memory functioning) and Hooper Visual Organization Test (HVOT – measures spatial skills).
Note: The authors did not report the actual r scores.

Known groups:
Clarke, Black, Badley, Lawrence & Williams, (1999) divided subjects at 3 months and 1 year post-stroke by level of impairment (mild-moderate-severe according to Adam’s Hemispheric Stroke Scale), by the presence or absence of depression (Zung Self-Rating Depression scale), by levels of physical disability (independent-moderately dependent-dependent according to the Functional Independence Measure), RNLI scores for these known groups demonstrated expected gradients and were significantly different as analyzed by analysis of variance. The difference in mean RLNI scores between categories in these analyses ranged from 12% to 62%.

Responsiveness

Wood-Dauphinee, Opzoomer, Williams, Marchand, and Spitzer (1988) administered the RNLI to a sample 70 patients to determine the responsiveness of the RNLI. They concluded that the scale is sensitive to change but the use of subscales provides a more accurate reflection as change (improvement or worsening) in specific domains could be hidden within the total score.

References

  • Bluvol, A., Ford-Gilboe, M. (2004). Hope, health work and quality of life in families of stroke survivors. Journal of Advanced Nursing, 48(4) 322-332.
  • Calmels, P., Pereira, A., Domenach, M., Pallot-Prades, B., Alexandre, C., Minaire, P. (1994). Functional ability and quality of life in rheumatoid arthritis: Evaluation using the Functional Independence Measure and the Reintegration to Normal Living Index. Revue Du Rhumatisme, 61(11), 723-731.
  • Clarke, P. A., Black, S. E., Badley, E. M., Lawrence, J. M., Williams, J. L. (1999). Handicap in stroke survivors. Disability and Rehabilitation, 21(3), 116-123.
  • Daneski, K., Coshall, C., Tilling, K., Wolfe, C.D.A. (2003). Reliability and validity of a postal version of the Reintegration to Normal Living Index, modified for use with stroke patients. Clinical Rehabilitation, 17, 835-839.
  • Daverat, P., Petit, H., Kemoun, G., Dartigues, J. F., Barat, M. (1995). The long term outcome in 149 patients with spinal cord injury. Paraplegia, 33, 665-668.
  • Dawson, D. R., Levine, B., Schwartz, M., Stuss, D. T. (2000). Quality of life following traumatic brain injury: A prospective study. Brain and Cognition, 44, 35-39.
  • Friedland, J. F., Dawson, D. R. (2001). Function after motor vehicle accidents: A prospective study of mild head injury and posttraumatic stress. The Journal of Nervous and Mental Disease, 189(7), 426-434.
  • Gauthier-Gagnon, C., Grise, M-C. (1994). Prosthetic Profile of the Amputee Questionnaire: Validity and reliability. Archives of Physical Medicine and Rehabilitation, 75, 1309-1314.
  • Gray, D. B., Hollingsworth, H. H., Stark, S. L., Morgan, K. A. (2006). Participation Survey/Mobility: Psychometric properties of a measure of participation for people with mobility impairments and limitations. Archives of Physical Medicine and Rehabilitation, 87(2), 189-197
  • Korner – Bitensky, N., Wood Dauphinee, S., Shapiro, S., Becker, R. (1994). Eliciting health status information by telephone after discharge from hospital: Health professionals versus trained lay persons. Canadian Journal of Rehabilitation, 8(1) 23-34.
  • Korner-Bitensky, N., Wood-Dauphinee, S., Siemiatycki, J., Shapiro, S., Becker, R. (1994). Health related information postdischarge: Telephone versus face-to-face interviewing. Archives of Physical Medicine and Rehabilitation, 75, 1287-1296.
  • May, L. A, Warren, S. (2002). Measuring quality of life of persons with spinal cord injury: external and structural validity. Spinal Cord, 40, 341-350.
  • Mayo, N. E., Wood-Dauphinee S., Cote, R., Gayton, D., Carlton, J., Buttery, J., Tamblyn, R. (2000). There is no place like home: An evaluation of early supported discharge for stroke. Stroke, 31, 1016-1023.
  • Mayo N., Wood-Dauphinee S., Cote R., Durcan L., Carlton J. (2002). Activity, participation & quality of life 6 months post-stroke. Archives of Physical Medicine & Rehabilitation, 83, 1035-1042.
  • McColl, M. A., Paterson, M., Davies, D., Doubt, L., Law, M. (2000). Validity and community utility of the Canadian Occupational Performance Measure. Canadian Journal of Occupational Therapy, 67(1), 22-33.
  • Patrick, L., Perugini, M. Leclerc, C. ( 2002). Neuropsychological assessment and competency for independent living among geriatric patients. Topics in Geriatric Rehabilitation, 14(4) 65-77.
  • Stark, D. L., Edwards, D. F., Hollingsworth, H., Grey, D. B. (2005).Validation of the Reintegration to Normal Living Index in a population of community-dwelling people with mobility limitations. Archives of Physical Medicine & Rehabilitation, 86(2), 344-345.
  • Steiner, A., Raube, K., Stuck, A. E., Aronow, H. U., Draper, D., Rubenstein, L. Z., Beck, J. C. (1996). Measuring psychosocial aspects of well-being in older community residents: Performance of four short scales. The Gerontologist, 36(1), 54-62.
  • Tooth, L.R., McKenna, KT., Smith, M., O’Rourke, P.K. (2003). Reliability of scores between stroke patients and significant others on the Reintegration to Normal Living (RNL) Index. Disability and Rehabilitation, 25(9), 433-440.
  • Trombly, C. A., Radomski, M. V., Davis, E. S. (1998). Achievement of self identified goals by adults with traumatic brain injury: Phase 1. The American Journal of Occupational Therapy, 52(10), 810-818.
  • Wood-Dauphinee, S. L., Opzoomer, M. A., Williams, J. I., Marchand, B., Spitzer, W. O. (1988). Assessment of global function: The Reintegration to Normal Living Index. Archives of Physical Medicine and Rehabilitation, 69, 583-590.
  • Wood-Dauphinee, S., Williams, J. I. (1987). Reintegration to normal living as a proxy to quality of life. Journal of Chronic Diseases, 40(6), 491-499.

See the measure

You can obtain the RNLI here.

Please click here to access the french language version. A research version is also available.

Table of contents

Screening for Self-Medication Safety Post-Stroke Scale (S-5)

Evidence Reviewed as of before: 07-01-2011
Author(s)*: Annabel McDermott, OT
Editor(s): Nicol Korner-Bitensky, PhD OT

Purpose

The Screening for Self-Medication Safety Post-Stroke Scale (S-5) is a screen for clinicians to identify patients’ self-medication safety and readiness following stroke. The tool can also be used by health professionals to make recommendations to improve self-medication skills of patients post-stroke (Kaizer, Kim, Van & Korner-Bitensky, 2010).

In-Depth Review

Purpose of the measure

The Screening for Self-Medication Safety Post-Stroke Scale (S-5) is a screen for clinicians to identify patients’ self-medication safety and readiness following stroke. (Kaizer, Kim, Van & Korner-Bitensky, 2010). It is a quick, inexpensive test that uses a checklist-style interview format.

Available versions

There is only one version of the Screening for Self-Medication Safety Post-Stroke Scale (S-5), which was developed by Kaizer, Kim, Van and Korner-Bitensky in 2010.

Features of the measure

Items of the measure:

The S-5 consists of 16 items that assess five domains:

  • Cognition (orientation; immediate and delayed memory recall)
  • Communication (comprehension; reading)
  • Motor function
  • Visual-perception
  • Judgement/executive functions/self-efficacy.

The patient must be able to correctly answer 2 of the first 3 questions regarding orientation to time and space in order to progress with the screen.

Scoring and Score Interpretation:

Each item is scored according to a yes/no response. There is no cumulative score. A score of “no” on any 1 item indicates the need for further assessment regarding this domain, or can be used to guide intervention planning to address this area of difficulty.

Each item also has a “concern” box, where the clinician can identify any concerns regarding the particular item. A summary “Concerns and Recommendations” section at the end of the tool also enables the clinician to document specific concerns and suggestions.

Equipment:

  • Pill bottle with childproof cap
  • Pill bottle without childproof cap
  • Pill bottle with a pharmacy label: must include the information commonly found on a label (medication name, dosage, frequency, time of day to take medication and the name of a person)
  • Liquid bottle with “push and turn” cover and a medicine cup
  • 1 syringe without needle
  • 8 disc-shaped white pills (e.g. shape of a vitamin C)
  • 1 oval-shaped blue or green gel-capsule pill
  • 1 oval shaped orange pill
  • 1 small and 1 larger disc-shaped white pill
  • Three objects: pen, coin & a key

Time:

The S-5 takes approximately 10 minutes to administer.

Training requirements:

No training requirements specified.

Subscales:

None.

Alternative forms of the S-5

Not applicable

Client suitability

Can be used with:

  • Clients following stroke.

Should not be used in:

  • Not specified.

In what languages is the measure available?

English.

Summary

What does the tool measure? Self-medication safety.
What types of clients can the tool be used for? Patients with stroke.
Is this a screening or assessment tool? Screening tool.
Time to administer Approximately 10 minutes.
Versions There are no alternative versions.
Other Languages There are no official translations.
Measurement Properties
Reliability Test-retest:
The test-retest reliability of the S-5 is currently under study.
Validity Content:
This tool is not intended as a comprehensive assessment of self-medication safety. Some daily self-medication tasks were intentionally not included due to its intended use as a screen only. Accordingly, content validity of this tool was reported as satisfactory.

Criterion:
Concurrent:
The concurrent validity has not been examined as there is currently no gold standard for assessing self-medication safety with this population.

Construct:
Known groups:
The known groups validity of the S-5 is currently under study.

Floor/Ceiling Effects Not yet examined.
Does the tool detect change in patients? Not yet examined.
Acceptability The S-5 is a quick and simple test to administer, with minimal equipment requirements and specific instructions for the assessor to follow.
Feasibility Administration of the S-5 is quick and easy, and can be performed by any member of the multidisciplinary team. Feedback from expert clinicians and patients indicates acceptable administration time, effort and complexity.
How to obtain the tool? Click here to see a copy of the S-5.

Psychometric Properties

Overview

Please refer to the article by Kaizer et al. (2010) for information regarding the psychometric properties of the S-5

References

See the measure

How to obtain the Assessment?

Click here to see a copy of the S-5

Table of contents

Stroke Impact Scale (SIS)

Evidence Reviewed as of before: 29-06-2018
Author(s)*: Lisa Zeltzer, MSc OT; Katherine Salter, BA; Annabel McDermott
Editor(s): Nicol Korner-Bitensky, PhD OT; Elissa Sitcoff, BA BSc

Purpose

The Stroke Impact Scale (SIS) is a stroke-specific, self-report, health status measure. It was designed to assess multidimensional stroke outcomes, including strength, hand function Activities of Daily Living / Instrumental Activities of Daily Living (ADL/IADL), mobility, communication, emotion, memory and thinking, and participation. The SIS can be used both in clinical and in research settings.

In-Depth Review

Purpose of the measure

The Stroke Impact Scale (SIS) is a stroke-specific, self-report, health status measure. It was designed to assess multidimensional stroke outcomes, including strength, hand function, Activities of Daily Living / Instrumental Activities of Daily Living (ADL/IADL), mobility, communication, emotion, memory and thinking, and participation. The SIS can be used both in clinical and research settings.

Available versions

The Stroke Impact Scale was developed at the Landon Center on Aging, University of Kansas Medical Center. The scale was first published as version 2.0 by Duncan, Wallace, Lai, Johnson, Embretson, and Laster in 1999. Version 2.0 of the SIS is comprised of 64 items in 8 domains (Strength, Hand function, Activities of Daily Living (ADL) / Instrumental ADL, Mobility, Communication, Emotion, Memory and thinking, Participation). Based on the results of a Rasch analysis process, 5 items were removed from version 2.0 to create the current version 3.0 (Duncan, Bode, Lai, & Perera, 2003b).

Features of the measure

Items:

The SIS version 3.0 includes 59 items and assesses 8 domains:

  • Strength – 4 items
  • Hand function – 5 items
  • ADL/IADL – 10 items
  • Mobility – 9 items
  • Communication – 7 items
  • Emotion – 9 items
  • Memory and thinking – 7 items
  • Participation/Role function – 8 items

An extra question on