CJS397127
CJS26110.1177/0829573510397127Hills et al.Canadian Journal of School Psychology
The School Short-Form
Coopersmith Self-Esteem
Inventory: Revised
and Improved
Canadian Journal of School Psychology
26(1) 62­–71
© 2011 SAGE Publications
Reprints and permission: http://www.
sagepub.com/journalsPermissions.nav
DOI: 10.1177/0829573510397127
http://cjs.sagepub.com
Peter R. Hills1, Leslie J. Francis2, and Penelope Jennings1
Abstract
The school short form of the Coopersmith Self-Esteem Inventory is a widely used
measure of children’s global self-esteem. Unlike the full-length scale, however, it
has been generally understood that the short form does not allow differentiation
between the major individual sources of self-esteem. The present study has examined
the internal structure of the school short form by exploratory and confirmatory
analysis on data provided by 3,056 adolescents between the ages of 13 and 15 years
and has demonstrated that after the removal of 6 redundant items, the newly revised
scale not only possesses improved psychometric properties but also contains three
clear factors that correspond to personal self-esteem and self-esteem derived from
parents and peers, respectively. The presentation of a revised version of the school
short form of the Coopersmith Self-Esteem Inventory that is psychometrically robust
and demonstrates three clear subscales will allow clearer distinctions to be made
among the sources of children’s self-esteem in future studies.
Résumé
La version abrégée, pour le milieu scolaire, de l’Inventaire d’estime de soi de Coopersmith
est grandement utilisée pour mesurer l’estime de soi globale chez les jeunes. On croit
généralement que cette version ne permet pas de différencier entre les principales
sources d’estime de soi, contrairement à ce qu’on obtient de la version complète. La
présente étude porte sur l’examen de la structure de la version abrégée, au moyen
d’analyses exploratoires et confirmatoires sur des données provenant de 3 056
adolescents, âgés de 13 à 15 ans. Or, il appert que la nouvelle échelle révisée, si on
1
St. Mary’s Centre, Wales, United Kingdom
University of Warwick, United Kingdom
2
Corresponding Author:
The Revd Canon Professor Leslie J Francis, University of Warwick, Warwick Religion and Education
Research Unit, Institute of Education, Coventry, CV4 7AL, United Kingdom
Email: [email protected]
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
63
Hills et al.
enlève six questions redondantes, possède non seulement de meilleures propriétés
psychométriques, mais comporte trois facteurs distincts qui correspondent à l’estime
de soi dérivée du soi, des parents et des pairs, dans l’ordre. C’est donc dire qu’une
version révisée de l’Inventaire d’estime de soi de Coopersmith pour le milieu scolaire,
qui comprend des propriétés psychométriques robustes et dégage aussi trois souséchelles claires, permettra de distinguer, à l’avenir, les sources d’estime de soi chez
les jeunes.
Keywords
children, confirmatory factor analysis, Coopersmith Self-esteem Inventory, positive
psychology, self-esteem
Self-esteem (SE) is a widely used construct both in popular and formal psychology
(Baumeister, Campbell, Krueger, & Vohs, 2003; Lipnevich, 2006). Maslow (1970)
identified satisfaction of the need for esteem as a contributory factor to positive
psychological functioning and proposed that the construct was made up of personal
SE and esteem that was generated by the positive regard of important others. SE has
been defined as an individual’s sense of self-worth, or the extent to which a person
values, approves of, appreciates, prizes, or likes him- or her-self (Blascovitch &
Tomaka, 1991), and has been related to many psychological domains, including
personality, behavior, socioeconomic factors, and health and clinical psychology.
The construct has also been widely used in educational psychology since the 1960s
(Coopersmith, 1967).
Many measures have been designed to assess SE, and by the mid-1970s several
authors (Drummond & McIntyre, 1977; Petersen, 1977; Wiley, 1974) commented
that these measures were generally inadequately supported by evidence of reliability or validity. One of the best-established instruments is the 50-item school form
of the Coopersmith Self-Esteem Inventory (SEI; Coopersmith, 1967, 1981) that
was devised for use with children and designed to assess attitudes toward the general self and in the specific contexts of school, parents, and peers. The factor structure of the 50-item instrument has been examined in several studies. Ketcham and
Morse (1965) identified five factors corresponding to total (personal?) self-esteem,
social self-esteem, doing well in school, self-deprecation, and self-certainty.
Kokenes (1978) identified nine factors, which were condensed into four bipolar
scales broadly related to school, parents, peers, and the general self and concluded
that the results supported the theoretical dimensions of the SEI, though 17 of the 50
items cross-loaded on at least 2 factors. Roberson and Miller (1986) extracted 8
empirical factors, though 13 items of the original scale were excluded because their
factor loadings on any of the factors were negligible. Correlations of the empirical
factors with the Coopersmith subscales gave most support for the existence of the
parental subscale, but the evidence for the existence of other subscales was
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
64
Canadian Journal of School Psychology 26(1)
ambiguous and one empirical factor did not load on any of the of the Coopersmith
subscales. The factor structure of the school form of the SEI does not appear to
have been investigated further in recent years.
The school short-form of the Coopersmith Self-Esteem Inventory was developed
to provide an alternative to the SEI when time for completion is limited. The scale
consists of the 25 items (from the 50-item scale) that showed the highest item–total
score correlations in the full scale. It was stated (Coopersmith, 1981) that this scale
does not allow differentiation by subscale and that the validity of the scale had not
been established. More recently, Zhang (1997) reported the internal reliability and
construct validity of the school short form to be satisfactory. A literature survey has
indicated that the short-form inventory is still in general use, mainly as a measure of
global self-esteem (e.g., Delaney & Lee, 1995; Francis, 2005; Francis & Gibbs, 1996;
Hills, Francis, & Jennings, 2006; Jones & Francis, 1996; Robbins, Francis, & Kerr,
2007; Sapp, 1994; Stark, Spirito, Lewis, & Hart, 1990; Williams, Francis, & Robbins,
2006), but the factor structure of the instrument appears not to have been investigated. The present study aims to establish whether the school short-form Coopersmith
Self-Esteem Inventory has an internal structure that might extend the usefulness of
the measure beyond its general application as a measure of global self-esteem.
Method
Participants
All 31 state-maintained secondary schools in Cornwall were invited to participate
in the project, and 23 accepted that invitation. Within the participating schools
questionnaires were administered during religious education lessons among Year 9
or Year 10 pupils (between the ages of 13 and 15 years). Pupils were assured of
anonymity and confidentiality. Completed questionnaires were received from
3,056 pupils (1,531 boys, 1,525 girls) of whom 34% were in Year 9 and 66% in
Year 10.
Measures
As part of a larger questionnaire concerned with attitudes toward religious education,
pupils completed the school short-form Coopersmith Self-Esteem Inventory (SEI;
Coopersmith, 1967), which consists of 25 items relating to three areas, to be answered
on a yes/no scale: (1) global self-esteem: “I can make up my mind without too much
trouble,” and “I often wish I were someone else”; (2) relations with parents, “My parents
usually consider my feelings,” and “My parents expect too much of me”; and (3) relations with peers, “I’m popular with kids [of] my own age,” and “Most people are
better liked than I am.” Self-esteem scores were calculated from the aggregate item
scores, with higher scores indicating greater self-esteem. A full listing of the items is
given in Table 1.
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
65
Hills et al.
Table 1. Exploratory Factor Analysis of the Coopersmith Self-Esteem Inventory Short Form
With Oblique Rotation
Item
M
F1
SD
C01 0.47 0.50 I often wish I were someone else (–)
0.65
C03 0.33 0.47 There are lots of things about myself I’d change
0.61
if I could (–)
C15 0.56 0.50 I have a low opinion of myself (–)
0.54
C17 0.70 0.46 I often feel upset in school (–)
0.47
C05 0.61 0.49 I get easily upset at home (–)
0.45
C13 0.59 0.49 Things are all mixed up in my life (–)
0.45
C23 0.67 0.47 I often get discouraged in school (–)
0.40
C12 0.54 0.50 It is pretty tough to be me (–)
0.39
C18 0.45 0.50 I am not as nice looking as most people (–)
0.38
C07 0.72 0.45 It takes me a long time to get used to anything new (–) —
C10 0.68 0.47 I give in very easily (–)
—
C02 0.45 0.50 I find it hard to talk in front of the class (–)
—
C04 0.69 0.46 I can make up my mind without too much trouble
—
C24 0.52 0.50 Things usually do not bother me
—
C25 0.66 0.48 I can’t be depended on (–)
—
C11 0.62 0.49 My parents expect too much of me (–)
—
C22 0.64 0.48 I usually feel as if my parents are pushing me (–)
—
C20 0.65 0.48 My parents understand me
—
C09 0.69 0.46 My parents usually consider my feelings
—
C16 0.51 0.50 There are many times when I would like to
—
leaveHome (–)
C08 0.75 0.43 I am popular with kids my own age
—
C06 0.85 0.36 I am a lot of fun to be with
—
C19 0.75 0.43 If I have something to say I usually say it
—
C14 0.46 0.50 Kids usually follow my ideas
—
C21 0.50 0.50 Most people are better liked than me (–)
—
Eigen value
5.34
26.1
Variance explained before rotation (%)
.74
Cronbach’s α
h2
F2
F3
—
—
— 0.38
— 0.29
—
—
—
—
—
—
—
—
—
—
—
—
—
-0.70
-0.66
-0.61
-0.58
-0.43
—
—
—
—
—
—
—
—
—
—
—
—
—
—
—
—
—
—
0.33
0.30
0.27
0.39
0.23
0.29
0.19
0.12
0.12
0.15
0.11
0.06
0.03
0.40
0.39
0.37
0.31
0.31
—
—
—
—
—
3.12
10.3
.76
0.64
0.56
0.41
0.38
0.36
1.34
6.9
.63
0.30
0.21
0.17
0.13
0.33
Note: h2 is the symbol for communality, and communality is the amount of variance in an item explained
by the factors.
Analysis Strategy
Exploratory factor analyses (EFAs) were conducted with the SPSS statistical Package
for Windows, Release 11.0.1 (SPSS for Windows, 2001). Confirmatory factor analyses (CFAs) were implemented with the AMOS structural equation modelling (SEM)
program (Arbuckle, 1997). Use of the chi-square statistic (χ2/df) is the most obvious
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
66
Canadian Journal of School Psychology 26(1)
way to measure the fit of a model to data, but unfortunately the χ2 goodness-of-fit
index is sensitive to sample size, and the probability of rejecting a hypothesized model
increases as the sample size increases. In consequence, many alternative goodness-of-fit
parameters have been devised to evaluate SEMs, but there is little agreement on those
that are the most useful, and it is now customary to report the results for a range of
indices. In addition to the chi-square statistic, model fit was tested in the present study
with the Goodness-of-Fit Index (GFI), the Tucker–Lewis Index (TLI), the ComparativeFit Index (CFI), the Normed-Fit Index (NFI), the Parsimony Normed Fit Index
(PNFI), and the Adjusted Goodness-of-Fit Index (AGFI). For most alternative measures of goodness of fit, as a rule of thumb, values in excess of 0.90 are considered to
indicate a good fit, though for the parsimonious indicators, PNFI and AGFI, values in
excess of 0.80 are considered acceptable (Hoyle, 1995).
Results and Discussions
Basic Statistics
Reported scores covered the theoretical range of the SEI (minimum = 0, maximum = 25)
and the average score (M = 15.03, SD = 5.33) was in excess of the theoretical midpoint
of the scale. In line with other studies (Francis, 1998; Primavera, Simon, & Primavera,
1974; Watkins, 1982), the mean scale score was greater for boys (M = 16.07, SD =
5.09) than for girls (M = 13.99, SD = 5.34) and independent t tests showed that the
difference was highly significant (t = 11.03, p < .001).
Internal Consistency
The internal consistency reliability (Cronbach’s α = .83) of the scale was adequate,
but examination of the individual item–rest-of-test correlations showed that the contributions made by several items were relatively small; for example, “Things usually
do not bother me,” and “I can’t be depended on,” had item–rest-of-test correlations
smaller than .20. The mean interitem correlation was .167, with individual values
ranging from –.009 to .577, which suggests that the scale is not homogeneous.
Exploratory Factor Analyses (EFA)
The suitability of the data for factor analysis was tested by calculating the Kaiser–
Meyer–Olkin measure of sampling adequacy, and the value obtained, 0.89, was more
than adequate (Kaiser, 1974). Principal components analysis extracted 6 factors with
eigen values >1, which together explained 48.1% of the total variance. The item compositions of the six factors following Varimax (orthogonal) rotation, not reported here,
identified several ambiguities that made it difficult to interpret the nature of the last
three factors that contained few items, and six items loaded more or less equally on
two factors. Orthogonal rotation produces uncorrelated factors, which may oversimplify the relationships existing in real-life data. A scree plot, however, suggested the
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
67
Hills et al.
presence of three factors. After extraction of three factors (maximum likelihood), the
data were accordingly processed by an oblique technique (direct oblimin, δ = 0),
which does not force the extracted factors to be uncorrelated (Table 1).
Table 1 shows the presence of three factors, and none of the items loaded on more
than one factor. Six items had factor loadings less than the applied cutoff value of 0.35;
all of these items appeared to be marginally associated with the first factor, had low
communalities ranging from 0.03 to 0.15, and were accordingly discarded. The three
factors were tentatively identified as personal self-esteem (F1), self-esteem derived
from parents (F2), and self-esteem derived from peers (F3). Of these, F1 explains substantially more of the total variance than F2 and F3.
After removal of the 6 items with weak factor loadings on any of the factors, the
revised scale comprised 19 items divided into 3 subscales: personal self-esteem (9 items),
parentally derived self-esteem (5 items), and peer-derived self-esteem (5 items). Given
the small number of items in each subscale, the scale reliabilities as measured by
Cronbach’s alpha and reported in Table 1 are considered sufficient according to the
criteria proposed by DeVellis (2003). As a measure of global self-esteem, the correspondence between the 19-item revised school short-form Coopersmith Self-Esteem
Inventory and its parent 25-item measure was high. The Pearson correlation between
the two scales was large (r = .97, p < .001). The scale reliability for the revised scale,
Cronbach’s α = .83, was unchanged despite the smaller number of items, and the
mean inter-item correlation also increased from .17 to .21, suggesting that the 6 discarded items are effectively redundant. The three subscales were intercorrelated
(F1 and F2, r = .56; F1 and F3, r = .60; and F2 and F3, r =.21), all significant at the
p <.001 level. These data demonstrate that both self-esteem derived from parents and
self-esteem derived from peers are closely related to personal self-esteem. However,
self-esteem derived from parents and self-esteem derived from peers are much less
closely related.
Confirmatory Factor Analysis (CFA)
A CFA analysis was first conducted on the 3-factor, 19-item model suggested by
the preceding CFA, and the results are reported in Table 2. The table shows that
both the GFI and then AGFI have values in excess of the recommended value of
0.90. Examination of the modification indices (MIs) suggested that the model
could be improved by allowing three pairs of error terms to covary. Covariance of
error terms is an indication that the items are statistically very similar in the
responses they generate, perhaps because they are semantically alike. The MIs
suggest that the addition of error-term covariances between Items C11 and C22
(“My parents expect too much of me,” and “I usually feel as if my parents are
pushing me”), Items C01 and C02 (“I often wish I were someone else,” and
“There are lots of things about myself that I would change if I could”), and Items
C17 and C23 (“I often feel upset in school,” and “I often get discouraged in
school”) would result in an improved goodness of fit. Since these items are clearly
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
68
Canadian Journal of School Psychology 26(1)
Table 2. Confirmatory Factor Analyses of the Revised Coopersmith Self-Esteem Inventory
Short Form
Goodness-of-fit indices
Absolute
2
Model
EFA model
Covariance between
Items C11 and
C22
Covariance between
Items C01 and
C03
Covariance between
Items C17 and
C23
χ
df
1919
1545
149
148
1337
147
1249
146
2
χ /df
Comparative
Parsimonious
GFI
TLI
CFI
NFI
PNFI
AGFI
0.933
0.945
0.844
0.876
0.864
0.892
0.854
0.883
0.744
0.764
0.915
0.930
9.09
0.953
0.893
0.908
0.898
0.772
0.939
8.56
0.956
0.901
0.915
0.905
0.773
0.942
12.9
10.4
Note: GFI = goodness-of-fit index; TLI = Tucker–Lewis index; CFI = comparative-fit index; NFI = normedfit index; PNFI = parsimony normed-fit index; AGFI = adjusted goodness-of-fit index.
semantically alike, the corresponding error-term covariances were added in turn.
The results that are also reported in Table 2 show that these modifications resulted
in progressive improvements and together provided values in excess of 0.90 for
six of the seven goodness-of-fit indices employed. Reference was made earlier to
the difficulties associated with the use of the χ2 goodness-of-fit index with large
samples, and with the present sample (N = 3,060), Table 2 shows that the
observed value, 8.56, is in excess of even the most relaxed recommended acceptable value, χ2/df < 5 (Wheaton, Muthén, Alwin, & Summers, 1977). When the
final analysis was repeated with a smaller randomly selected sample (n = 250),
the χ2/df value fell to 1.82 less than the most conservative recommended acceptable value of 2 (Byrne, 1989). It was, therefore, concluded that the revised school
short-form Coopersmith Self-Esteem Inventory has a psychometrically sound
factor structure.
Conclusions
The present study set out to examine the internal structure of the school short form
of the Coopersmith Self-Esteem Inventory, a widely used measure of children’s
global self-esteem, and also to explore ways of developing and improving this
well-established measure. It has been shown that after the removal of six redundant items, the newly revised instrument possesses improved psychometric properties and also contains three clear factors. Three main conclusions emerge from
the study.
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
69
Hills et al.
First, the correlation between the original 25-item form of the instrument and the
new 19-item scale is high, r = .97, so the present study does not call into question the
findings of several decades of research that have employed the former 25-item instrument as a measure of global self-esteem.
Second, one of the main reasons for the development of the 25-item form of the SEI
was to provide a shorter and more economical instrument for research purposes when
the time for administration of the questionnaire was limited. The present findings suggest, however, that there remains unnecessary redundancy in the 25-item instrument
and that the development of an even shorter form has produced not only a more economical scale but has also resulted in a scale with enhanced psychometric properties.
Third, a supposed disadvantage of the 25-item short form in comparison with the
50-item parent instrument is that it has been generally understood that the short form
does not allow differentiation among the major sources of self-esteem (Coopersmith,
1981). The present findings, however, have demonstrated that the revised scale contains three clear factors corresponding to personal self-esteem and self-esteem derived
from parent and peers, respectively.
On the basis of these conclusions three practical recommendations may be
advanced. The first recommendation concerns the use of this revision of the school
short-form of the Coopersmith Self-Esteem Inventory in further research. This
instrument has the advantage of clear continuity, with established research using
both the 50-item form and the 25-item form, and the additional advantages being it
is shorter and quicker to complete (with only 19 items), is psychometrically robust
without any redundancy, and distinguishes between three sources of self-esteem.
The second recommendation concerns the potential offered for reanalysis of existing data generated by studies using the 25-item form. Since all the constituent items of
the revised scale are present in the original scale, it will be relatively easy to adapt
previously collected data in order to isolate and to explore the correlates of the three
different sources of self-esteem.
The third recommendation concerns the potential application of this 19-item instrument among other screening instruments and assessment batteries in contexts of psychological and counseling practice. In a relatively unobtrusive way the factor structure
of this short instrument may be useful in providing insight not only into general levels
of self-esteem but also into the sources of self-esteem, particularly as these sources
affect relationships with parents or relationships with peers.
Author’s Note
Sadly Dr. Peter R. Hills died shortly after the completion of this manuscript. His coauthors wish
to dedicate the paper in memory of collaboration and friendship.
Declaration of Conflicting Interests
The authors declared no potential conflicts of interests with respect to the authorship and/or
publication of this article.
Funding
The authors received no financial support for the research and/or authorship of this article.
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
70
Canadian Journal of School Psychology 26(1)
References
Arbuckle, J. L. (1997). Amos users guide version 4. Chicago: Smallwaters Corporation.
Baumeister, R. F., Campbell, J. D., Krueger, I. K., & Vohs, K.D. (2003). Does high selfesteem cause better performance, interpersonal success, happiness, or healthier lifestyles?
Psychological Science in the Public Interest, 4, 1-44.
Blascovitch, J., & Tomaka, J. (1991). Measures of self-esteem. In J. P Robinson, P. R. Shaver, &
L. S. Wrightsman (Eds.), Measures of personality and social psychological attitudes (Vol. 1).
San Diego, CA: Academic Press.
Byrne, B. M. (1989). A primer of LISREL: Basic applications and programming for confirmatory
factor analytic models. New York/Berlin: Springer-Verlag.
Coopersmith, S. (1967). The antecedents of self-esteem. Palo Alto, CA: Consulting Psychologists Press.
Coopersmith, S. (1981). Self-esteem inventories. Palo Alto, CA: Consulting Psychologists Press.
Delaney, W., & Lee, C. (1995). Self-esteem and sex roles among male and female high school
students: Their relationship to physical activity. Australian Psychologist, 30, 84-87.
DeVellis, R. F. (2003). Scale development: Theory and applications. London: Sage.
Drummond, R. J., & McIntyre, W. G. (1977). Evaluating the factor structure of “self-concept” in
children: A cautionary note. Measurement and Evaluation in Guidance, 9, 172-176.
Francis, L. J. (1998). Is there gender bias in the short form Coopersmith Self-esteem Inventory?
Educational Research, 40, 83-89.
Francis, L. J. (2005). God images and self—esteem: A study among 11— to 18—year olds,
Research in the Social Scientific Study of Religion, 16, 105-121.
Francis, L. J., & Gibbs, D. (1996). Prayer and self-esteem among 8- to 11-year-olds in the
United Kingdom. Journal of Social Psychology, 136, 791-793.
Hills, P., Francis, L. J., & Jennings, P. (2006). Religious behaviour, personality and dimensions of
self-esteem among 13- to 15-year-old adolescents. Journal of Research on Christian Education, 15, 61-76.
Hoyle, R. H. (1995). The structural equation modelling approach: Basic concepts and fundamental issues. In R. H. Hoyle (Ed.), Structural equation modelling, concepts, issues and
applications. Thousand Oaks, CA: Sage.
Jones, S. H., & Francis, L. J. (1996). Religiosity and self-esteem during childhood and adolescence. In L. J. Francis, W. K. Kay, & W. S. Campbell (Eds.), Research in religious education
(pp. 189-205). Leominster, MA: Gracewing.
Kaiser, H. F. (1974). An index of factorial simplicity, Psychometrika, 39, 31-36.
Ketcham, W., & Morse. W.C. (1965). Dimensions of children’s social and psychological development related to school achievement (Cooperative Research Project No. 1286, Office of
Education, Department of Health). Ann Arbor: University of Michigan.
Kokenes, B. (1978) A factor analytic study of the Coopersmith Self-Esteem Inventory. Adolescence, 13, 149-155.
Lipnevich, A. A. (2006). Low self-esteem: Myth or reality? Focus on Basics, 8, 1-7.
Maslow, A. H. (1970). Motivation and personality. New York: Harper & Row.
Petersen, A. C. (1977). The measurement of self among adolescents. Journal of Youth and Adolescence, 6, 201-203.
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016
71
Hills et al.
Primavera, L. H., Simon, W. E., & Primavera, A. M. (1974). The relationship between self-esteem
and academic achievement: An investigation of sex differences. Psychology in Schools,
11, 213-216.
Robbins, M., Francis, L. J., & Kerr, S. (2007). God images and self-esteem among secondary
school pupils in South Africa. In P. Heggy (Ed.), What do we imagine God to be? The function
of “God images” in our lives (pp. 89-107). Lampeter, UK: Edwin Mellen.
Roberson, T. G., & Miller, E. (1986). The Coopersmith Self-esteem Inventory: A factor analytic
study. Educational and Psychological Measurement, 46, 269-273.
Sapp, M. (1994). Cognitive-behavioral counseling: Applications for African-American middle
school students who are academically at-risk. Journal of Instructional Psychology, 21, 161-171.
SPSS for Windows. (2001). Chicago: SPSS Inc.
Stark, L. J., Spirito, A., Lewis, A. V., & Hart, K. J. (1990). Encopresis: Behavioral parameters associated with children who fail medical management. Child Psychiatry and Human
Development, 20, 169-179.
Watkins, D. (1982). Sex-role perceptions of Filipino adolescents. International Journal of Psychology, 17, 359-368.
Wheaton, B., Muthén, B., Alwin, D. F., & Summers, G. F. (1977). Assessing reliability and
stability in panel models. In D. R. Heise (Ed.), Sociological methodology. San Francisco:
Jossey-Bass.
Wiley, R. C. (1974). The self-concept. Vol. 1: A review of methodological considerations and
measuring instruments. Lincoln: University of Nebraska Press.
Williams, E., Francis, L. J, & Robbins, M. (2006). Rejection of Christianity and self-esteem.
North American Journal of Psychology, 8, 193-195.
Zhang, W. (1997). Primary study of characteristics of self-esteem in junior high school students.
Psychological Science (China), 20, 504-508.
Bios
Peter R. Hills, PhD, following his retirement held honorary research fellowships in the
University of Wales, Bangor, and the St. Mary’s Centre, Wales, United Kingdom. Sadly he died
before seeing the publication of this article.
Leslie J. Francis, PhD, ScD, DD, is professor of religions and education within the Warwick
Religions and Education Research Unit, University of Warwick, England, United Kingdom.
Penelope Jennings, PhD, undertook her doctoral research in the University of Wales, Bangor, and
continues her association with the research group at St Mary’s Centre, Wales, United Kingdom.
Downloaded from cjs.sagepub.com at PENNSYLVANIA STATE UNIV on September 12, 2016