Advice on total-score reliability issues in psychosomatic measurement
Klaas Sijtsma⁎, Wilco H.M. Emons
Tilburg University, Tilburg, The Netherlands
Received 24 May 2010; received in revised form 2 November 2010; accepted 4 November 2010
Objective: This article addresses three reliability issues that
are problematic in the construction of scales intended for use
in psychosomatic research, illustrates how these problems may
lead to errors, and suggests solutions. Methods: We used
psychometric results and present five computational studies.
The first, third, and fourth studies are based on the generation
of artificial data from psychometric models in combination with
distributions for scale scores, as is common in psychometric
research, whereas the second and fifth studies are analytical.
Results: The power of Student's t test depends more on
sample size than on total-score reliability, but reliability must
be high when one estimates correlations involving test scores.
Short scales often do not allow total scores to be significantly
different from a cutoff score. Coefficient alpha is uninformative
about the factorial structure of questionnaires and is one of the
weakest estimators of total-score reliability. Conclusions: The
relationship between questionnaire length/reliability and statis-
tical power is complex. Both in research and individual
diagnostics, we recommend the use of highly reliable scales so
as to reduce the chance of faulty decisions. The conclusion
calls for profound statistical research producing hands-on rules
for researchers to act upon. Factor analysis should be used to
assess the internal consistency of questionnaires. As a
reliability estimator, alpha should be replaced by better and
readily available methods.
© 2010 Elsevier Inc. All rights reserved.
Keywords: Coefficient alpha; Individual decision making; Power analysis; Short scales
This article addresses three important but often neglected
issues related to total-score reliability of scales intended for
use in psychosomatic and other health-related research.
These scales measure not only, for example, depression and
anxiety [1,2], chronic fatigue , and Type D personality 
but also coping, emotions, and compliance and health-
related quality-of-life aspects, such as mobility around the
house after surgery  and individual feelings and percep-
tions about people's daily life .
The first issue is that researchers too easily accept
minimum values for total-score reliability in research
settings. We show that such values depend on the research
goal and discuss the statistical power of Student's t test and
estimation of correlations as examples. We draw some
tentative conclusions and notice that a finer-grained advice
requires a larger-scale statistical investigation.
The second issue is that researchers use short and
reliable scales for research and individual decision making
more and more. We argue that even for satisfactory total-
score reliability, person measurement using short scales
may be imprecise, evoking many decision errors . We
provide some research results but noticed that this topic
also requires large-scale research before definitive advice
can be given.
The third issue concerns coefficient alpha , which is
often used both as an index of internal consistency,
meaning the degree to which the items in a questionnaire
measure the same attribute, and as an estimator of total-
score reliability. We argue that, contrary to common belief,
alpha is not an index of internal consistency and further
that, as a reliability estimator, it should be replaced by
available superior alternatives.
Journal of Psychosomatic Research 70 (2011) 565–572
⁎Corresponding author. Department of Methodology and Statistics,
FSW, Tilburg University, P.O. Box 90153, 5000 LE Tilburg, The
Netherlands. Tel.: +31 13 4663222 or +31 13 4662544; fax: +31 13
E-mail address: email@example.com (K. Sijtsma).
0022-3999/10/$ – see front matter © 2010 Elsevier Inc. All rights reserved.
Our purpose is to make the reader aware of these three
highly complex issues and the risks involved when they are
underestimated in practical scale construction. Underestima-
tion happens when one assumes that (a) a total-score
reliability of, say, .6–.8 is sufficient for testing hypotheses
about mean scores without ascertaining whether statistical
power is high enough; (b) a scale consisting of a small
number of high-quality items is reliable enough for
individual decision making without checking whether
enough decisions are correct; and (c) an alpha value of,
say, .8 means that the items measure the same attribute
without investigating their factorial structure.
This article is technical, but we tried to get our points
across using only a few equations. The first two issues are
too complex to lead to definitive practical guidelines, and we
noticed that more profound statistical research than we can
provide here is needed for deriving such guidelines. The
third issue concerning coefficient alpha has led to clear-cut
conclusions resulting in unambiguous advice. We hope this
article stimulates researchers to think hard about their scales,
the methodological problems they must resolve, and the
psychometric methods they use.
Reliability requirements in research
Measurement instruments may be used in psychosomatic
research or for assessing patients' individual behavior.
Research applications address, for example, the comparison
of means between groups of patients and controls and
estimating correlations of total scores with treatment
outcomes. Individual assessment refers to the measurement
of an individual with the purpose of deciding, for example,
whether he or she should receive treatment or whether his or
her health condition has improved due to treatment.
Researchers and practitioners have to use their resources
efficiently, and a reasonable question is whether short
scales may be used instead of longer scales. A longer scale
has higher total-score reliability, but perhaps a short scale
based on only the highest-quality items leads to little loss
of statistical power and estimation precision (research) and
reliability (assessing individual performance). In this
section, we discuss the consequences of using short scales
in research and individual assessment. These constitute the
first two issues we cited. First, we discuss the classical test
CTT, total-score reliability
We assume a questionnaire contains J items and each
item has an index number j running from 1 to J and scores
running from 0 to m. The total score on the J items is the sum
of the item scores, Xj, and defined as X=∑j=1
questionnaire has six rating-scale items with scores running
from 0 through 4, then respondents can have total scores, X,
running from 0 to 24. The scale length (SL) is the range of
possible X scores and equals mJ; here, SL=24.
JXj. If the
CTT splits total score X in a systematic part or true
score T and a random measurement error E, such that
X=T+E . Total-score reliability quantifies the degree
to which performance in a group is systematic and
predictable as opposed to being random and unpredict-
able. Reliability is defined in two ways, which are
First, suppose it were possible to readminister the
questionnaire a second time under exactly the same
circumstances as the first. The reliability of the total score
is then technically defined as the correlation between the
total scores obtained at the first and second occasions,
denoted X and X′, and the reliability is the correlation, ρXX′.
This definition answers the question that is pervasive in
statistics: What would happen if I could do it again?
Second, in a group, the total-score variance can be
decomposed into the sum of the true-score variance and the
random error variance, σX
the proportion of total-score variance that is due to true-score
larger share of true-score variance. If it were possible to
readminister the questionnaire a second time under exactly
the same circumstances as the first, the reliability would
remain the same: rel=rel′. The first definition is mathemat-
ically identical with the second, so that ρXX′=σT
Thus, the degree to which total scores can be repeated under
the same circumstances equals the ratio of true-score
variance to total-score variance in both administrations.
reliability assumes values between 0 and 1.
An individual's true score T differs for different
instruments measuring the same attribute. Assume we have
a four-item questionnaire in which each item has a rating
scale with scores 0,…,4, so that total score X has values
0,…,16. Let John have a true score TJohn=6 and Mary a true
score TMary=10. We double the length of the questionnaire
by another four-item set, which is parallel with the first four-
item set (parallelism is: the two item sets are psychometri-
cally identical but have different contents ); then, the true
scores also double: TJohn=12 and TMary=20, and so does their
difference. In real life, subsets of items are never parallel and
true scores change without neatly doubling.
2. Reliability is defined as
2). Higher reliability means a
2and because variance is nonnegative, the
Research instrument, reliability, and sample size
Testing group-means difference
We considered the following research question: Suppose
a group of patients and a group of controls have different
mean depression levels. A researcher has constructed a
research instrument meant for measuring depression. He or
she intends to compare the two groups using the instrument.
For this purpose, the researcher has collected data from both
groups and used Student's t test to test whether the mean
instrument scores in the two groups are equal (null
hypothesis) or not. How does the total-score reliability affect
the statistical test results? This is a complex problem (e.g.,
566K. Sijtsma, W.H.M. Emons / Journal of Psychosomatic Research 70 (2011) 565–572
Refs.  and ) to which no easy answers are available,
but here we provide tentative suggestions.
In general, as reliability increases, the share of the true-
score variance in the total-score variance is larger and one
expects group differences in mean true scores to become
larger. Hence, one also expects differences between mean
total scores to become larger. Thus, a more reliable
questionnaire is expected to better pick up existing
differences in depression levels between groups, and
Student's t test may signal a smaller difference already as
significant. What makes this problem complex is that
reliability may increase in different ways: by keeping total-
score variance constant while letting true-score variance
increase, by keeping true-score variance constant while
letting error variance decrease, by letting the number of items
grow by adding items of the same psychometric quality, by
starting with a small set of the best items and then adding
items of lower quality, and so on. The point is that the
manipulation of many questionnaire and item properties
affects reliability and it is rather unpredictable how Student's
t test reacts to this manipulation. We briefly consider the
power of Student's t test for increasing number of items (J),
also called questionnaire length.
Student's t statistic for independent samples with unequal
sizes and variances equals (P=patients, C=controls,
t =XP− XC
Increasing the questionnaire length, and hence the
reliability, affects the group means and the S.E. values in
the t statistic, but it is difficult to predict what happens to
the t test's power.
We conducted a computational study using simulated
data for five-point rating-scale items to investigate the power
of the t test as questionnaire length increases and hence
reliability increases. Because data are simulated, the
alternative hypothesis is known (details about the study
can be obtained from the second author). Table 1 shows that
as questionnaire length increased from 5 to 20 items,
Cronbach's alpha increased from .70 to .90. The difference
between the group means (numerator of t statistic) and the
corresponding S.E. (denominator) both increased, and
statistic t and standardized effect size d increased but only
little. P values decreased, and at the 5% level, we found a
significant result for J=15 and J=20 but not for J=5 and
J=10. The power of the t test increased little as questionnaire
length increased but greatly as sample size increased for
fixed reliability (Table 1, three rightmost columns). We
conclude that for testing group-means difference, more
power is gained by increasing sample size than by
increasing reliability, say, from .80 to .90. Does this result
hold for any statistical analysis? The answer is no, as the
next example shows.
Estimating correlations with other variables
Suppose one wishes to know the correlation between
the depression total score X and a criterion score Y
indicating suitability for a particular treatment, which is
also represented by a fallible score. Then, to learn how the
reliabilities of both measurements affect their correlation,
ρXY, the attenuation correction formula  from CTT can
be written as
with TX and TY denoting the true-score components of
total scores X and Y, respectively.
Clearly, the correlation between the true scores is of
interest, but inpractice, one onlyhas accesstothe fallibletotal
scores. The equation shows that ρXY is attenuated by
measurement error: the lower the reliabilities, the lower the
correlation between the observable scores and the more ρXYis
a distorted estimate of ρTXTY. Arbitrarily, assume that ρTXTY=.6,
which is the correlation of interest; then, only if ρXX′=ρYY′=1
does one find that ρXY=ρTXTY. In practice, perfect reliability is
too good to be true. Lowering one or both reliabilities yields a
between 0 and 1; hence, ρXYis always
lower than ρTXTY. For example, take ρXX′=ρYY′=.8, then
ρXY=.8×.6=.48. For ρXX′=.6 and ρYY′=.8, one finds
:48 × :6c:42.
We conclude that the lower the reliabilities, the lower
the correlation between the two total scores and hence
the more distorted the correlation of interest. A larger
sample does not help; the same distortion is estimated
but with more precision. The effect of larger sample size
on the precision is illustrated using another computa-
Table 2 shows for ρTXTY=.6 and varying combinations of
reliabilities of X and Y the resulting correlation ρXY.
We took each of the 10 population values of ρXY as
correlations found in real research (i.e., as sample
values rXY), and then for increasing sample size N, we
computed confidence intervals (CIs) for the population
correlation ρXY. For this purpose, we used the Fisher
Questionnaire length and Student's t-test results
Shown for four levels of questionnaire length (J) are the coefficient
alpha, difference in means (X---
statistic, P value, and standardized effect size (d) for N=100 and small
effect size, and the power of independent samples t test for varying
sample sizes (N) for small effect size and a nominal significance level of
5%. For each value of J, 1000 data sets were simulated. Table entries are
means across 1000 data sets.
C), S.E. (denominator t statistic), t
567K. Sijtsma, W.H.M. Emons / Journal of Psychosomatic Research 70 (2011) 565–572
z-transformation (writing r instead of rXYand ln for the
2ln1 + r
to normalize the sampling distribution of r. The distribution
of z has S.E.,
S:E: z ð Þ =
N − 3
Using S.E.(z), we computed 95% CIs, z±1.96S.E.(z),
and then used the inverse transformation (writing exp for
the exponential function),
Þ − 1
Þ + 1;
to transform the lower and upper bounds of the interval
back to the scale of r. This produced the well-known
asymmetrical CIs for the correlation ρXY shown in
Table 2 shows that for each correlation (third column), the
way the CIs are estimated maintains the same midpoint based
on the correlation (but not identical with it, due to
asymmetry) but produces a narrower interval as sample
size N grows. Hence, by using the sample correlation and the
sample size, one estimates the same distorted parameter ρXY
but with more precision.
We do not recommend using the formula for attenuation
correction the other way around by inserting the reliabilities
and the correlation between total scores X and Y and then
estimating the correlation between the true scores. This
estimate is based on fallible total scores, and the question
on what would really happen with the correlation between
X and Y when both measures were improved by adding
more items or replacing malfunctioning items by better
items remains unanswered. The attenuation correction gives
a prediction, not a value based on real instrument
construction and real data collected by means of these
To summarize, when group means are compared, a higher
reliability increases statistical power only little and a larger
sample size is more effective. When the correlation of a total
score with another total score is estimated, imperfect
reliability distorts the desired outcome. Fleshing out the
minimum-reliability problem for all kinds of hypotheses and
estimation problems requires profound statistical research,
which is beyond the scope of this study. Based on the present
state of knowledge, rules of thumb for total-score reliability
are difficult to set up: Sometimes reliability seems to be
subordinate to sample size, and sometimes it is dominant,
nearly independent of sample size; hence, the general
problem is complex and in need of further research.
Reliability requirements in individual assessment
For determining the usefulness of total scores in making
decisions about individuals, size N of the sample in which
the psychometric qualities of the scale were ascertained does
not play a role. What does play a role are the individual's J
item scores. The greater J, the more information is available
about the individual's true scores and the more precise is the
individual's true-score estimate. In this section, we discuss
first the precision of total score X, then the use of total score
X for classifying patients into one of two clinical groups, and
finally the use of total-score change for assessing individual
change due to therapy.
We use the total score X to estimate T. Let the estimated
value be denoted by T̂, then T̂=X. If John has a score of 19
points, we take this to be his true-score estimate. This is what
almost all applied researchers do—that is, they use X, but
more sophisticated methods would lead to the same
conclusion. It is of great interest to know how precise the
estimate T̂=X is. For this purpose, psychometricians use the
CI for T. Very few researchers use CIs and instead take X as
if it were the true score, tacitly assuming ρXX′=1; but then
why bother about total-score reliability at all?
To determine a CI, we need a statistical model for
drawing samples—in this case, a total score from a
distribution of total scores obtained with the same instru-
ment. Ideally, for each individual, such a distribution would
be available, but in real life, each individual produces only
one total score. In the absence of such individual distribu-
tions, the common practice is to use the S.D. of the
measurement error for the whole group, σE, and assume it
also is the S.D. of an individual's hypothetical distribution of
total scores. S.D. σEis called the standard measurement error
(SME), and equals,
The mean of an individual's hypothetical distribution of total
scores is his or her true score; for John, that is μJohn=TJohn.
SME = rE= rX
Correlation between scale scores X and Y (ρXY) and corresponding 95% CIs
for 10 combinations of reliability of X (ρXX′) and Y (ρYY′) and five levels of
N, given a true-score correlation (ρTXTY) of .6
−.04 to .49
568 K. Sijtsma, W.H.M. Emons / Journal of Psychosomatic Research 70 (2011) 565–572
Furthermore, it is assumed that John's distribution is
normal; in statistical notation, XJohn∼N(TJohn, σE
SME is estimated by inserting the S.D. of the total score in
the group, SX, which estimates σX, and an estimate rXX′
for the total-score reliability ρXX′, for example, Cronbach's
alpha, so that,
A (1−α)×100% CI is computed as T̂John±z1/2αSE, where
z1/2αis the normal deviate that corresponds to the area under
the normal curve for the CI. For a 95% CI, in a table for the
standard normal distribution, one finds z0.025=1.96; for a
90% CI, one finds z0.050=1.645. With the use of SEand a
choice for z1/2α, a CI with the same width can be computed
for each true-score estimate.
Individual assessment, reliability, and number of items
Classification using the total score
Suppose one uses total score X for assigning people to
one of two groups. The decision rule is: if XbXc(where c
is cutoff), assign to the no treatment condition, and if
X≥Xc, assign to the treatment condition. Then, one has to
check whether the cutoff falls in the CI for a particular
individual. If it does, then total score X is not significantly
different from the cutoff and a reliable decision cannot be
made; if it does not, then lower total scores support the
nontreatment decision and higher total scores support the
In practice, reliability is often used in a somewhat
ritualistic way. For example, if reliability is, say, at least .8, it
is assumed that the scale is precise enough for making
individual decisions, and total scores are used as if they were
true scores. However, .8 is smaller than 1, and as long as a
scale has imperfect reliability, one should test whether the
true score equals the cutoff. Next, we demonstrate how
reliability is related to SMEs, CIs, and SL. This yields
surprising results, urging researchers and practitioners to be
cautious, especially when they use short scales.
We did an experiment using the item response theory
model known as the graded response model  to define
the psychometric properties of a set of J items comprising a
questionnaire and combined the artificial item set with a
distribution of the attribute scores in a group of interest.
Next, we generated data for N individuals who responded to
J items with five ordered answer categories yielding scores
0,…,4, analyzed the data set, and repeated the experiment for
different numbers of items and varying item and question-
naire properties. Items had their highest discrimination
power around cutoff Xc, and total-score reliability for the
longest questionnaire equaled .9. Shorter scales consisted of
item subsets from the longer questionnaire. (Please consult
the first author for more information.)
Table 3 shows total-score reliability, SME, 90% and
95% CIs for true score T, which is located exactly at half
the scale, and the ratio of CI and SL (CI/SL). The question
is whether true-score estimate T̂differs significantly from
cutoff Xc, which lies to the right of T at 60% of the scale.
Only if T̂is significantly smaller than Xcis the individual
admitted to nontreatment. Table 3 shows (from bottom to
top) that reliability and SME decreased as questionnaire
length J decreased. Nevertheless, many researchers would
consider the reliability sufficient for short scales, and their
view seems to be supported by smaller SMEs resulting in
narrower CIs; hence, it seems that the true score can be
estimated with greater precision. However, given the
smaller number of items and the corresponding loss of
information, this cannot be true. The catch is that SL has
decreased faster than the CIs, so that CI/SL at least has
doubled. Fig. 1 shows that the cutoff has moved into the
CI for small J and that it can no longer be concluded that
the individual should be assigned to nontreatment.
The example shows that, in spite of good reliability
based on good-quality items, short tests can deceive in a
terrible way. Emons et al.  have taken this topic further
and demonstrated that small numbers of good-quality
items easily classify respondents in the wrong condition.
The problem is small SL, which leaves total scores less
room to be different from cutoffs even when CIs are
narrow. Long(er) scales reduce percentages of classifica-
tion errors considerably.
Assessing individual change
We considered effect of therapy as reflected in total-score
change. Individual change is the difference between an
individual's total score after treatment, X2, and his or her
total score before treatment, X1. The reliable change (RC)
index  is defined as follows:
RC =X2− X1
where Sdiff =
hypothesis of no change. The use of the RC statistic assumes
normally distributed measurement errors. One may test (e.g.,
at the 5% level) whether observed change in the expected
direction of recovery (X2−X1N0) is significant (RC N1.645)
or whether observed change in any direction, recovery or
deterioration (X2−X1≠0), is significant (∣RC∣N1.96).
For the same computational setup that underlay the results
in Table 3, Table 4 shows power results for improvement
(SEis the sample SME) under the hull
SL, coefficient alpha, SME, 90% and 95% CIs for the true score (T) half way
the scale, and CI/SL for each CI for four levels of questionnaire length (J)
Results for the whole table are based on a data set of 1000 simulated item-
score vectors using five-point rating-scale items (i.e., m=4).
569 K. Sijtsma, W.H.M. Emons / Journal of Psychosomatic Research 70 (2011) 565–572
(one-tailed test) and change (two-tailed test). First, in
identical circumstances, the one-tailed test always has
more power than the two-tailed test. Second, power is only
substantial (say, N.60) if the effect sizes are large (1 S.D. of
the true person-distribution) or very large (2 S.D.s). For
small effect size (0.25 S.D.), even 20-item tests fail to
produce a powerful one-sided test (power=.37); for medium
effect size (0.5 S.D.), only tests in excess of 10 items have
Again, questionnaire length proves to be important in
individual decision making. This should come as no surprise
as change assessment also involves classification: Did the
individual improve or change significantly or not? Jacobson
and Truax  rightly argued that in addition to statistically
significant change, an individual must show clinically
significant change in order to be recovered. This requires
the individual to cross a cutoff score Xc that can be
determined in different ways depending on the dysfunctional
and functional total-score distributions (Ref. , p. 13).
Jacobson and Truax (, p. 16) also suggested statistically
testing whether a posttreatment score X2 is significantly
greater than cutoff Xc. In our computational study, we only
considered the power of the RC statistic, but if we would also
require individuals having significant RC values to show
clinical significance and to have posttreatment scores
significantly larger than Xc, the power of the whole
procedure would be smaller compared with the results in
Table 4; also see Jacobson and Truax  for a discussion of
Coefficient alpha: internal consistency and
Coefficient alpha is often used as an index for internal
consistency. The literature isnot explicitabout the meaning of
the term, butinternalconsistency isusually taken to mean that
the items constituting the questionnaire measure the same
attribute [14,15]. A high value of alpha is then interpreted to
mean that the items measure this attribute. However, several
authors [15–17] have argued convincingly that there is no
underlying the item scores. Thus, they argued that both a one-
Fig. 1. Ninety percent CIs for a true score located at 50% of the scale, for different questionnaire lengths, and a cutoff score that is at 60% of the scale.
Power to detect treatment effects using Jacobson and Truax’ RC index and significance level of 5% for four levels of individual clinical change
Power of one-tailed test Power of two-tailed test
Individual changeIndividual change
Small MediumLarge Very largeSmall MediumLarge Very large
Results in each cell are based on 50,000 newly simulated difference scores.
570 K. Sijtsma, W.H.M. Emons / Journal of Psychosomatic Research 70 (2011) 565–572
have either low or high alpha values. Rather than repeating
their results, we call attention to the following line of
reasoning, which makes the same point.
The information about the factorial composition of a set of
items comprising a questionnaire is in the structure of the
correlations between the items. For convenience, we use the
inter-item covariances instead of the inter-item correlations.
Roughly speaking, if all inter-item covariances are the same
(Table 5, left panel), one factor can describe this inter-item
items within the same cluster covary highly and items from
different clusters covary lowly (Table 5, right panel), then two
factors can describe this two-cluster inter-item covariance
structure, and so on. Now, if alpha were an index for internal
consistency, meaning that one factor explains the inter-item
covariance structure (i.e., the items measure the same
attribute), only one-factorial inter-item covariance structures
and a look at the equation for alpha explains why.
Let the inter-item covariance between items j and k be
denoted σjk, and let the mean value of the inter-item
covariances between all J(J−1) item pairs (in Table 5, 12
pairs) in the questionnaire be denoted σ̄, then we can write
coefficient alpha as follows:
Thus, alpha depends on the mean of the inter-item
covariances but not on the individual inter-item covariances.
Since the information about the factorial structure of the J
items is in the structure of the J(J−1) inter-item covariances
and since this information islost when onlythe mean of these
covariances is available, alpha cannot provide information
about the factorial item structure. For example, for both
examples in Table 4, one finds that σ̄ =.06 (and that J=4 and
about the factorial structure and is misleading as an index of
The second use of alpha, as an estimator of total-score
reliability, is correct but can be somewhat unfortunate, as we
explain next. Researchers often do not seem to realize that
alpha is only a lower bound to the reliability—that is,
alpha≤ρXX′(e.g., Ref. ), and that much better estimators
are available. Sijtsma  recommends using Guttman's λ2
2=1.72), so that alpha=.56. Hence, alpha is uninformative
, which, similar to alpha, is available in IBM SPSS
Statistics 18, or the greatest lower bound (GLB; ), which
is the lower bound closest to reliability ρXX′. It is worth
noticing that alpha≤λ2≤GLB≤ρXX′. For example, for an
eight-item questionnaire on coping strategies, Sijtsma 
reported alpha=.778, λ2=.785, and GLB=.852. GLB is
available in EQS (http://www.mvsoft.com/eqs60.htm; e.g.,
One could argue that all a lower bound, such as alpha,
does wrong is underestimate total-score reliability and that a
conservative reliability estimate may even spur researchers
to construct a questionnaire that is more reliable than they
think. On the other hand, using a reliability estimate that
improves upon alpha provides a more realistic reliability
estimate and produces narrower CIs for individual decision
making. Using better reliability estimates alleviates the need
for longer scales at least a little.
Alpha, λ2, and GLB can be estimated irrespective of the
factorial structure of the items; they simply are not
informative about it. However, researchers usually want
their questionnaires to measure one attribute. Thus, it is
advisable to investigate the factorial structure of the items
and perhaps remove one or two items that are poor indicators
of the dominant factor prior to estimating total-score
reliability for the remaining items. Structural equation
modeling also provides methods for reliability estimation
that may yield better results than alpha, provided the
underlying factor model describes the data well . This
factor model may contain, for example, second-order factors
in addition to a dominant factor.
Conclusions and recommendations
We conclude by providing a couple of take-home
messages for instrument constructors, researchers using
instruments, and practitioners:
• Higher reliability produces more power for testing
differences between means, but a larger sample is more
effective. Also, higher reliability reduces the underes-
timation of the strength of relationships, but sample size
has no effect. More research into similar power and
estimation problems is badly needed. In general, we
recommend using longer scales having a highly reliable
• Statistically, the number of items in a scale is the
sample of information available for estimating indivi-
duals' attribute levels. Short scales tend to result in
faulty decision making. Our advice is to use many items
if the decision is important.
• Longer scales that include the items from the short-
scale version do everything technically better than the
short scale, but they increase the burden on the patient
and require more administration time. Alternatively, the
researcher may consider using many sources of
Hypothetical covariance matrices for one-factorial questionnaire data and
two-factorial questionnaire data, both with alpha=.56
One-factorial questionnaire Two-factorial questionnaire
571 K. Sijtsma, W.H.M. Emons / Journal of Psychosomatic Research 70 (2011) 565–572
information, among them several short scales for Download full-text
different attributes (e.g., Ref. ).
• Coefficient alpha is not an index for internal consis-
tency of the questionnaire. Internal consistency must be
investigated using methods for dimensionality assess-
ment, such as factor analysis.
• Coefficient alpha should be replaced by better reliabil-
ity estimation methods, such as λ2and the GLB.
We thank Marrie H. J. Bekker and Susanne S. Pedersen
for providing critical comments on a previous draft of
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