The Gerontologist
Vol. 43, No. 1, 121–127
Copyright 2003 by The Gerontological Society of America
Psychometric Properties of an Abridged
Version of the Zarit Burden Interview Within
a Representative Canadian Caregiver Sample
Norm O’Rourke, PhD,1 and Holly A. Tuokko, PhD2
Purpose: Given the exponential increase in dementia
prevalence anticipated in the coming years, measurement of caregiver burden has become common in
gerontological research and clinical practice. The
Zarit Burden Interview (BI) has emerged as the most
widely utilized burden measure. The current study
examines the psychometric properties of responses to
an abridged, 12-item version of this scale. Design
and Methods: Data were derived from a national
epidemiological study of dementia incidence and
patterns of care (N 5 1,095). Informal caregivers of
surviving institutionalized and community-dwelling
index subjects were interviewed 5 years subsequent
to initial recruitment (n 5 770). Results: Results of
both the exploratory and confirmatory factor analyses
support a two-factor structure of responses to this
abridged scale. Subsequent to control for demographic variables, dementia illness features, and baseline depressive symptoms at baseline, responses to
this brief BI provide a significant increase to prediction
of depressive symptoms at Time 2 (R2 5 .24, p ,
.01) with no additional variance provided by the 10
remaining items from the complete BI (R2 5 0, ns).
Implications: The results of this study are discussed
relative to theory and the operational definition of
caregiver burden. Findings can be generalized with
The Canadian Study of Health and Aging (CSHA) was funded by the
Seniors’ Independence Research Program through the National Health
Research and Development Program (NHRDP) of Health Canada
(Project 6606-3954-MC [S]). Additional funding was provided by Pfizer
Canada Incorporated through the Medical Research Council/Pharmaceutical Manufacturers Association of Canada Health Activity Program,
the NHRDP (Project 6603-1417-302 [R]), Bayer Incorporated, and the
British Columbia Health Research Foundation (Projects 38 [93-2] and 34
[96-1]). The CSHA was coordinated through the University of Ottawa
and the Division of Aging and Seniors, Health Canada. We thank all
staff across Canada involved in the CSHA.
Address correspondence to Norm O’Rourke, PhD, Gerontology
Research Centre, Simon Fraser University at Harbour Centre, #2800–515
West Hastings Street, Vancouver, British Columbia, Canada V6B 5K3.
E-mail: [email protected]
1
Gerontology Research Centre and Programs, Simon Fraser University at Harbour Centre, Vancouver, British Columbia, Canada.
2
Centre on Aging and Department of Psychology, University of
Victoria, Victoria, British Columbia, Canada.
Vol. 43, No. 1, 2003
121
greater confidence given the representative and
national composition of caregivers recruited for this
study.
Key Words: Burden, Caregivers, Dementia, Test
Reliability, Test validity
With awareness of the demands faced by those
who provide care to persons with dementia,
measurement of caregiver burden has become
common in gerontological research and clinical
practice. This construct has been defined as a context-specific negative affective outcome occurring as
a result of one’s perceived inability to contend with
role demands (O’Rourke, Haverkamp, Tuokko,
Hayden, & Beattie, 1996). Implicit in this definition,
various factors are believed to moderate caregivers’
ability to cope over time (cf. O’Rourke, Haverkamp,
Rae, et al., 1996; Russo, Vitaliano, Brewer, Katon, &
Becker, 1995).
Despite some conceptual ambiguity (Braithwaite,
1992), burden is commonly understood within the
context of the cognitive phenomenological model of
stress and coping (Lazarus & Folkman, 1984).
Burden is thus seen to emerge as a function of one’s
subjective belief that current and future resources are
insufficient to meet role demands (Zarit, 1990). As
a result, negative secondary appraisal (or burden) is
believed to function as a harbinger of problematic
physical and mental health outcomes, such as
depressive symptomatology (O’Rourke & Tuokko,
2000). Support is found in the literature for this
operational definition of caregiver burden (e.g.,
Clyburn, Stones, Hadjistavropoulos, & Tuokko,
2000; Lutzky & Knight, 1994).
The 22-item Burden Interview (BI; Zarit, Orr, &
Zarit, 1985) has emerged as the most commonly used
measure of caregiver burden. In recent years,
however, abridged versions of the BI have been
devised with the assumption that burden will be
assessed more consistently if a reliable and valid
brief measure can be identified (cf. Hébert, Bravo, &
Table 1. Descriptive Features and Psychometric Properties of
Study Variables
Instrument
M
SD
a Kurtosis Skewness
Activities of Daily Living 13.11 9.61 .96
Dementia Behavior
15.72 12.80 .86
Disturbance
CES-D, Time 1
7.45 8.49 .88
CES-D, Time 2
7.68 9.67 .91
Full Burden Inventory
15.58 13.76 .90
Brief Burden Inventory
7.56 7.84 .85
21.34
1.44
.15
1.08
1.60
1.81
1.20
1.85
2.59
3.45
1.18
1.37
Notes: The distribution of CES-D scores at Time 2 indicates
positive skewness outside of normal parameters (Tabachnick &
Fidell, 2001). Recalculation of CES-D scores to attain a more
normal distribution (i.e., square root transformation,
skewness 5 2.50) did not significantly alter study findings.
CES-D 5 Center for Epidemiologic Studies–Depression scale.
Préville, 2000; Knight, Fox, & Chou, 2000; Whitlatch, Zarit, & von Eye, 1991). One such scale has
recently been proposed by Bédard and colleagues
(2001). The 12 items of this brief BI were selected as
those with the highest item-total correlations.
Similar to the original measure, these authors
contend that responses to their brief BI reflect two
distinct factors (personal strain and role strain) with
acceptable indices of internal consistency for both
(i.e., a 5 .88 and a 5 .78, respectively). On the basis
of various analyses, Bédard and colleagues (2001)
conclude that their scale remains an effective
measure of caregiver burden despite its relative
brevity.
The current study provides a more rigorous
examination of the psychometric properties and
factor structure of this brief BI. As opposed to
participant data derived from one circumscribed
geographic region (Bédard et al., 2001), caregivers
for the current study were randomly recruited from
each Canadian province. In addition, longitudinal
data were examined to ascertain the predictive
validity of this brief BI relative to depressive
symptoms as measured 5 years subsequent to initial
measurement.
Methods
Participants
Care Recipients.—Older adults in the community
and institutional settings were recruited as part of
a national epidemiological study of dementia prevalence in Canada. The methodology used by the
Canadian Study of Health and Aging (CSHA) is
described in detail elsewhere (CSHA Working
Group, 1994a). In brief, persons over 64 years of
age were randomly identified from governmental
health records in all provinces (except Ontario,
where enumeration records were used). A total of
9,008 community-dwelling older adults underwent
clinical screening during which the Modified MiniMental State Examination (3MS; Teng & Chui,
122
1987) was administered. As compared with Folstein’s
original Mini-Mental State Examination, the 3MS
provides greater gradation of scores and covers
a broader range of cognitive abilities (e.g., abstract
reasoning, generative naming). Scores on the 3MS
range from 0 to 100, with lower totals suggestive of
cognitive impairment (Tombaugh, McDowell, Kristjansson, & Hubley, 1996).
Community-dwelling participants scoring below
78/100 on the 3MS and all those in institutions were
invited to undergo clinical examination (N 5 2,928).
On the basis of all clinically relevant information,
a consensus diagnosis was reached by interdisciplinary teams composed of a physician, neuropsychologist, nurse, and/or psychometrician (Tuokko,
Kristjansson, & Miller, 1995).
Caregivers.—Primary caregivers were identified
by index subjects and/or their families as the person
most responsible for day-to-day decisions. A primary
caregiver was identified in all but seven cases (CSHA
Working Group, 1994b). Of this total, 32 caregivers
(2.8%) could not be contacted and 38 (3.4%)
declined participation. Interviews were conducted
in a location selected by the caregiver, at which time
a trained research assistant administered all study
measures (see Table 1 for descriptive statistics).
Interviews were conducted in the home of the
caregiver (77.9%), index subject (4.6%), or another
location selected by the caregiver (e.g., workplace;
17.5%). This flexibility was based on prior findings
that those unable to travel to medical facilities
represent a distinct subset of caregivers (Dura &
Kiecolt-Glaser, 1990).
The CSHA caregiver sample included those
supporting index subjects in the community (n 5
469, Time 1), as well as those residing in institutions
(n 5 626, Time 1). This decision to include both
groups was based, in part, on prior findings suggesting that neither involvement nor dysphoria appears to
decline markedly for caregivers once their relative has
been placed in institutional care (see Dura, Stukenberg, & Kiecolt-Glaser, 1991; Russo et al., 1995).
Participation in the current study, however, was
restricted to family or friends of persons with
dementia (i.e., exclusion of paid or formal caregivers),
because care provision is believed to be inherently
more distressing for those personally committed to
care recipients (N 5 1,095, Time 1). It is assumed that
paid caregivers represent a distinct population.
A follow-up study was undertaken to examine
patterns of change in cognition and well-being of
older adults, as well as the health and welfare of
those who provide care (CSHA Working Group,
2002). All but 33 caregivers from the initial wave of
data collection were located (5 years, 2 days on
average after initial participation, SD 5 3 months,
10 days). For analysis of predictive validity of the
brief BI, participation was restricted to caregivers of
index subjects who remained alive (n 5 790, Time
The Gerontologist
2). Caregivers of those who had died were excluded
from this set of analyses, because it was assumed
that bereavement would confound the relationship
between burden and depressive symptoms.
Measures
Care Recipient Illness Characteristics.—Activities of Daily Living (ADLs). Caregivers were asked
to rate care-recipients’ ability to perform 14 separate
ADLs (both self-maintenance and instrumental
activities). This measure was taken from the Older
Americans Resources and Services Questionnaire
(Multidimensional Functional Assessment of Older
Adults; Fillenbaum, 1988). Respondents were asked
to rate each ADL along a 3-point Likert-type scale
[without help (0), some help (1), unable to do
unaided (2)]. The range of possible ADL scores is 0–
28, with higher totals reflecting greater impairment.
Dementia Behavior Disturbance (DBD) Scale.
Caregivers were administered a separate measure to
assess problematic behaviors directly related to
neurodegenerative illness. The DBD scale (Baumgarten, Becker, & Gauthier, 1990) consists of 28
questions. Caregivers are asked to rate how frequently each behavior had occurred in the past week
(e.g., ‘‘Hoards things for no obvious reason’’;
‘‘Makes unwarranted accusations’’). Responses are
reported along a 5-point Likert-type scale [never (0)
to all of the time (4)]. Possible scores range from 0 to
112, with higher totals reflecting greater disturbance.
In contrast to previous measures, the DBD scale
was developed with a more narrow definition of
behavioral disturbance to focus on the specific
manifestations of dementia syndromes. Items, therefore, do not tap functional and somatic symptoms nor
cognitive impairment as Baumgarten and colleagues
(1990) contend that these features are related to illness
severity, as opposed to behavioral disturbance.
From the initial validation study, responses to the
DBD scale appear to possess optimal internal
consistency (a 5 .83). Test-retest reliability over
a 2-week period was reported as r 5 .71 (Baumgarten et al., 1990). Construct validity has been
established relative to the Behavior and Mood
Disturbance Scale, as responses to these two
measures are strongly correlated (r 5 .73). These
findings suggest that responses to the DBD provide
a valid and reliable index of problematic behaviors
attributable to neurodegenerative illness.
Caregiver Variables.—Demographic information
was obtained at the time of initial participation.
These questions included the age of caregivers, years
of formal education, marital status, and relationship
to the index subject. Caregivers were also asked
whether or not they had experienced a series of health
problems over the past year (i.e., allergies, chest
problems, heart condition, kidney disease, cancer,
diabetes, high blood pressure, arthritis/rheumatism,
Vol. 43, No. 1, 2003
123
digestive troubles, nervousness, stroke, and insomnia). A cumulative variable was computed on the
basis of endorsement of these health conditions.
Center for Epidemiologic Studies–Depression
Scale (CES-D). The CES-D (Radloff, 1977) was
administered at both baseline and follow-up. This
measure asks respondents to rate the frequency of
various depressive symptoms over the past week
(e.g., ‘‘I had trouble keeping my mind on what I was
doing’’; ‘‘I felt lonely’’). Responses are provided
along a 4-point Likert-type scale ranging from rarely
or none of the time (0) to most or all of the time (3).
Among general adult samples, a cutoff score of 16
has been identified as suggestive of clinical depression (Radloff & Teri, 1986). However, a CES-D
cutoff score above 20 has been proposed for older
adults, given the prevalence of chronic health
conditions that may be mislabeled as depression
symptoms (e.g., insomnia, loss of energy; Himmelfarb & Murrell, 1983; Lyness et al., 1997).
Lewinsohn, Seeley, Roberts, and Allen (1997)
indicate that the utility of the CES-D is not
compromised by age, gender, physical disease, or
cognitive or physical impairment. Although developed and first validated with general adult populations, the CES-D appears appropriate for use with
older adults (Lewinsohn et al., 1997).
The BI. The BI (Zarit, Orr, & Zarit, 1985) presents
caregivers with a series of 22 questions regarding perceived strain in caring for persons with dementia. The
degree to which caregivers endorse each item is rated
along 5-point Likert-type scales. The range of possible
BI scores is 0–88, with higher totals reflecting greater
burden. Reported reliability coefficients for responses
to the full-scale range from a 5 .88 (Hassinger, 1986)
to a 5 .94 (O’Rourke & Wenaus, 1998).
The brief BI as proposed by Bédard and colleagues
(2001) is composed of 12 of 22 BI items (see Table 2).
Similar to the original scale, these authors contend
that responses to this brief BI reflect two underlying
constructs (i.e., personal strain and role strain).
Concurrent validity of responses to the abridged
scale was examined relative to indices of patient
behavioral disturbance and ADL impairment (Bédard et al., 2001).
Results
Data were missing at both points of measurement
(estimated at roughly 5% of usable data from CSHA-I
and CSHA-II). The PRELIS program was used to
estimate values for missing data (Jöreskog & Sörbom,
1996b). PRELIS imputes values on the basis of like
responses as opposed to substituting mean item
scores. According to Little and Rubin (1987), this
method is preferable to substitution with mean values
as the latter can obscure between group differences.
Consistent with prior research, the majority of
caregivers in this study were female (780/1,095 or
71.2%). Adult children composed the majority of
Table 2. Abridged Zarit Burden Interview Items and Factor Composition as Proposed by Bédard and Colleagues (2001)
Item
1.
2.
3.
4.
5.
6.
7.
8.
9.
10.
11.
12.
Factor
Do you feel that because of the time you spend with ___ that you do not have enough time for yourself?
Do you feel stressed between caring for ___ and trying to meet other responsibilities for your family or work?
Do you feel angry when you are around ___?
Do you feel that ___ currently affects your relationship with other family members or friends in a negative way?
Do you feel strained when you are around ___?
Do you feel your health has suffered because of your involvement with ___?
Do you feel that you don’t have as much privacy as you would like because of ___?
Do you feel that your social life has suffered because you are caring for ___?
Do you feel you have lost control of your life since ___’s condition?
Do you feel uncertain about what to do about ___?
Do you feel you should be doing more for ___?
Do you feel you could do a better job in caring for ___?
(timeself, PS)
(stressed, PS)
(angry, PS)
(relation, PS)
(strained, PS)
(health, PS)
(privacy, PS)
(soclife, PS)
(control, PS)
(uncertain, RS)
(domore, RS)
(dobetter, RS)
Notes: Items appear as adapted for the Canadian Study of Health and Aging (CSHA Working Group, 1994b). Abbreviated
item names correspond to those shown in Figure 1. PS 5 personal strain factor; RS 5 role strain factor.
caregivers (56.7%) with a significant percentage of
spouses (21.2%). The remainder self-identified as
either friends or members of the index subjects’ extended family (e.g., daughters-in-law, grandchildren,
nieces, and cousins). Given this percentage from
younger generations, it is not surprising that the
average age of caregivers (M 5 58.64 years, SD 5
12.65) was notably lower than index subjects (M 5
85.53 years, SD 5 6.50).
Response levels to this brief BI were significantly
higher for female (M 5 8.29, SD 5 8.22) versus
male caregivers [M 5 5.77, SD 5 6.47; t(1,093) 5
4.87, p , .01]. Also, internal consistency as
measured by Cronbach’s alpha is somewhat higher
for women (a 5 .85) than men (a 5 .82). Combined,
a 5 .85, which indicates no appreciable loss in
internal consistency, compared with the complete
measure (a 5 .90). For both the full scale and brief
BI, internal consistency of responses falls within
optimal parameters (i.e., .90 a .80; Clark &
Watson, 1995; DeVellis, 1991).
Factor Structure of Responses to the Brief BI
Two hundred participants were randomly selected
from the overall sample at Time 1 to examine the
factor structure of responses to the brief BI.
Exploratory factor analysis (EFA) was performed
with maximum-likelihood extraction and varimax
rotation. The Kaiser-Meyer-Olkin (KMO) measure
of sampling adequacy indicated sufficient interrelatedness among items for factor analysis (KaiserMeyer-Olkin 5 .84).
Initial estimates suggested a three-factor solution
based on the Kaiser-Guttman criterion (i.e., 3
eigenvalues greater than 1.0), with only two providing more than 10% of observed variance [k1 5
4.41 (36.7%), k2 5 1.78 (14.8%), k3 5 1.06
(8.76%)]. As noted by Floyd and Widaman (1995),
the Kaiser-Guttman criterion generally provides
overinclusive factor solutions. For instance, the
Cattell-Nelson-Gorsuch scree test suggested a twofactor solution as visual inspection of the scree plot
124
indicated a leveling of eigenvalues after Factor 2. On
the basis of this observation, a two-factor EFA was
computed next.
Consistent with the results reported by Bédard
and colleagues (2001), each of the 12 items
significantly loaded on their respective factors (9 on
personal strain and 3 on role strain); however, 4 of
12 items loaded significantly across factors (i.e.,
estimates .30). For the item, ‘‘Do you feel
uncertain about what to do ...?,’’ the loading
estimate on personal strain (.40) was greater than
that on role strain (.30) in contrast to factor structure
reported by Bédard and colleagues (2001).
Confirmatory factor analysis (CFA) was performed next on remaining Time 1 data to further
examine the viability of this two-factor structure
(n 5 895). CFA was computed with LISREL
software (LInear Structural RELations 8; Jöreskog
& Sörbom, 1996a). This a priori solution emerged as
viable, because all items loaded significantly on their
respective factors (i.e., t values >j1.96j). Similar to
EFA findings, the ‘‘uncertain about what to do’’ item
significantly loaded across factors (see Figure 1).
Subsequent to correction for correlated error
between item pairs, the CFA model indicated
effective fit of data [v2(df 5 36) 5 79.72, p , .01].
For instance, the Comparative Fit Index (CFI) is
within optimal parameters (i.e., CFI . .95; CFI 5
.99) as are the Adjusted Goodness-of-Fit Index
(AGFI; i.e., AGFI .90; AGFI 5 .97) and the Root
Mean Square Error of Approximation (RMSEA; i.e.,
RMSEA , .05; RMSEA 5 .037; see Hu & Bentler,
1999). These results support the viability of the twofactor structure of responses to the brief BI as
proposed by Bédard and colleagues (2001).
Predictive Validity of Responses to the Brief BI
Hierarchical regression analysis was next computed to ascertain the predictive validity of brief BI scores
relative to depressive symptoms, as measured 5 years
subsequent to initial administration of the BI (i.e.,
CES-D scores as the dependent variable). Baseline
The Gerontologist
Figure 1. Confirmatory factor analysis model for two-factor brief Burden Interview (n 5 895). Maximum likelihood estimates
(standardized solution and significance levels). Asterisks denote parameters fixed to 1.0 for scaling and statistical identification. As
a result, significance levels cannot be computed for these two items.
descriptive variables (age, sex, years of formal
education, marital status, relationship to care recipient) were first entered to control for demographic
effects on depressive symptoms (R2 5 .14, p , .01).
Dementia illness characteristics at Time 1 were
entered next as covariates (inability to perform ADLs,
dementia-related behaviors, cognitive impairment;
R2 5 .02, p , .01). CES-D totals at baseline were
entered to control for initial depressive symptoms
(R2 5 .08, p , .01). The brief BI was next entered
again leading to a significant increase in prediction of
CES-D scores at Time 2 (R2 5 .02, p , .01).
Residual items from the original BI were entered last.
This final step was undertaken to determine any
loss in predictive validity as a result of item
reduction. This would not appear to be the case,
because no change in the predictive strength of this
regression equation was observed subsequent to
inclusion of the 10 items discarded from the
complete scale (R2 5 0, ns). This suggests no loss
of predictive validity for the brief BI relative to the
original 22-item scale. These results are notable
given the extended duration between measurement
of both caregiver burden and CES-D scores and
rigorous control for demographic variables, dementia illness features, and depressive symptoms at
baseline (see Table 3).
Consistent with the stress and coping paradigm,
regression analyses presuppose that burden is an
antecedent of depressive symptomatology. In other
words, caregiver burden is believed to co-occur or
precede the emergence of depressive symptoms (cf.
Lawton, Moss, Kleban, Glicksman, & Rovine, 1991;
Vitaliano, Russo, Young, Teri, & Maiuro, 1991).
Vol. 43, No. 1, 2003
125
Analyses reported herein support this interpretation,
because burden appears to predict significantly
depressive symptomatology over and above rigorous
control for baseline variables.
Of note, reversal of variables in an attempt to
ascertain the predictive validity of depressive symptomatology at Time 1 relative to caregiver burden at
Time 2 failed to lead to a significant increase in burden
variance (subsequent to control for caregiver demographics, dementia illness variables, and baseline
depressive symptoms; R2 5 .003, ns). This finding
provides further support for the hypothesized causal
association in which burden is assumed to precede or
co-occur with depressive symptomatology.
Cutoff Criteria for the Brief BI
According to Bédard and colleagues (2001), a score
above 16 on the brief BI is suggestive of clinically
significant caregiver burden. This threshold was
identified on the basis of response score distributions. The current study examined this brief BI
cutoff score vis-à-vis depressive symptoms (both
measured at baseline). Relative to a cutoff score of
20 on the CES-D, responses to the brief BI exhibited
a specificity rate of 90% (i.e., exclusion of 9/10
caregivers providing concurrent responses below
threshold on the CES-D). At 49%, the level of
sensitivity for this brief BI cutoff score was
considerably less optimal. This latter result suggests
that a score above 16 on the brief BI identified less
than half of caregivers presenting within the
clinically significant range on the CES-D. A cutoff
Table 3. Hierarchical Regression Analysis of Caregiver
Depressive Symptoms at Time 2 (n 5 770)
B
SE B
b
F
Age of caregiver
Caregiver sex
Relationship to care recipient
Marital status
Years of education
Chronic health conditions
.02
.21
2.15
.19
2.20
.70
.03
.72
.08
.25
.10
.21
.03
.01
2.06
.03
2.07
.13
.63
.09
3.61
.58
4.15*
11.27**
Activities of daily living
Dementia-related behaviors
Index subject cognitive status
2.03
2.04
.15
.05
.03
.34
2.03
2.01
.02
.50
.02
.18
Depressive symptoms at
baseline
.29
.05
.26
38.15**
Brief Burden Interview
.29
.07
.24
17.35**
Residual burden items
2.09
.08
2.06
Variable
1.40
2
Notes: R 5 .14 for inclusion of caregiver demographics
(p , .01); R2 5 .02 subsequent to inclusion of patient illness
variables (p , .01); R2 5 .08 subsequent to inclusion of
brief Burden Interview (p , .01); R2 5 .02 subsequent to inclusion of depressive symptoms at baseline (CES-D, Time 1;
p , .01); R2 5 0 subsequent to inclusion of residual Burden
Interview items (ns). CES-D 5 Center for Epidemiologic Studies–Depression scale.
score above 10 increases sensitivity to 75% relative
to the CES-D; however, specificity drops to 68%.
These findings suggest that the cutoff score
proposed by Bédard and colleagues (2001) does not
serve as an effective index within this caregiver
sample. It would therefore appear premature to
propose a definitive cutoff. Similar to research and
clinical applications with the original measure, the
strength of this brief BI would appear to be as
a continuous measure as opposed to a screening
instrument.
Discussion
The results of this study provide general support
for the psychometric properties of responses to the
Bédard brief BI. Consistent with initial findings,
internal consistency of responses to this measure
remains within optimal parameters. Findings from
the current study also support the factor structure
reported by Bédard and colleagues (2001). Although
EFA findings suggest that four items load across
factors, CFA results support the two-factor structure
with only one cross-loaded item.
Also of note, this brief BI appears to possess
predictive validity relative to depressive symptomatology measured 5 years subsequent. Despite rigorous control for demographic variables, dementia
illness variables and depressive symptoms at baseline,
brief BI scores provide a significant increase in the
predictive strength of this regression equation. In
contrast, those items deleted from the original BI
provide no incremental increase in observed variance.
126
However, the cutoff score proposed by Bédard
and colleagues (2001) does not appear to provide an
effective index of depressive symptomatology. Scores
below the suggested cutoff identified the vast
majority of asymptomatic caregivers, yet the sensitivity of this response level is less than adequate.
This would indicate that this threshold is not viable,
because identification of clinically significant dysphoria is more critical than exclusion of those who
present as euthymic.
A further finding of note pertains to the negligible
contribution of dementia illness variables relative to
dysphoria among caregivers. Although the grouping
of illness variables contributed significantly to prediction of depressive symptomatology, none of the
variables attained individual significance. By contrast, the physical health of caregivers and years of
education provided unique and incremental increase
in prediction of depressive symptoms. As noted by
Baumgarten (1989), caregiver characteristics are
strongly associated with role-related distress in
contrast to patient variables (e.g., ADL impairment,
dementia-related behaviors, and illness severity).
Results of the current study extend this observation
from analysis of cross-sectional data to prediction of
dysphoria several years in the future. One interpretation of this finding is that illness-related factors
may trigger depressive symptomatology, yet the
extent of symptoms is expressed relative to individual caregiver variables (O’Rourke, Haverkamp,
Tuokko et al., 1996). In other words, the severity of
depressive symptoms is determined as a function of
caregiver or dyad-specific characteristics, such as the
personality of caregivers (Hooker, Monahan, Shifren, & Hutchinson, 1992), prior psychiatric history
(Russo et al., 1995), and quality of the premorbid
relationship (Williamson, Shaffer, & The Family
Relationships in Late Life Project, 2001).
Results of the current study are more compelling
because of the composition and size of the sample
examined. In contrast to those recruited by Bédard
and colleagues (2001), CSHA participants represent
a randomly derived national sample. The findings of
this study can thus be generalized with greater
confidence. However, various methodological limitations of the CSHA should be noted. For instance, all
questionnaires were administered in a serial order
raising the possibility of carryover and order effects.
Because questionnaires were not counterbalanced,
these potential confounds can be neither measured nor
discounted. Also of note, caregiver data were obtained
via face-to-face interviews. Without provision for
anonymity, responses to scales such as the BI may have
been influenced by response biases (e.g., socially
desirable responding; O’Rourke & Wenaus, 1998).
It should also be noted that caregivers in both
studies responded to the complete set of BI items.
Results may have differed had just the brief BI been
administered as opposed to these selected items
embedded within the complete 22-item scale. Future
The Gerontologist
research should administer just these 12 items when
attempting to replicate current findings. Also of note,
predictive validity was assessed on the same
responses as included in item analyses (because of
sample size considerations). Ideally, it would have
been preferable to perform calculations on separate
participant subsets, as repeated use of data may limit
the ability to generalize current findings.
Overall, this study provides substantial support for
the psychometric properties of the Bédard brief BI. No
loss of reliability or validity would appear to result
from reduction of items from the original scale.
Consistent with the findings previously reported by
Bédard and colleagues (2001), this brief BI appears to
be an effective measure of caregiver burden.
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Received January 24, 2002
Accepted July 22, 2002
Decision Editor: Laurence G. Branch, PhD