J Gerontol B Psychol Sci Soc Sci-1998-Ranzijn-P96-P104

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Journal of Gerontology: PSYCHOLOGICAL SCIENCES

1998, Vol. 53B, No. 2, P96-P104

Copyright 1998 by The Gerontological Society of America

The Role of Self-Perceived Usefulness and Competencein the Self-Esteem of Elderly Adults:

Confirmatory Factor Analyses of the BachmanRevision of Rosenberg's Self-Esteem Scale

Rob R anzijn, John Keeves, Mary Luszcz, and N . T. Feather

School of Psychology, The Flinders University of South Australia, Adelaide.

This article reports on a confirmatory analytic study of the Bachma n revision (1970) of Rosenberg's Self-Esteem

Scale (1965) that was used in the Australian L ongitudinal Study of Ageing (ALSA). Participants comprised 1,087

elderly people a ged between 70 and 10 3 years (mean 77 years). Five competing factor models w ere tested with

LISREL 8. The best-fitting m odel was a nested one, with a General Self-esteem second-order factor and two first-

order factors, Positive Self-regard and Usefulness/Comp etence. This model was validated with data from a later

wave of ALSA. Usefulness and c ompetenc e hav e received little attention in the gerontological literature to date. Pre-

liminary results indicate that usefulness/competence may be an important predictor of well-being. Further w ork is

required on the relationships among usefulness, competence, self-esteem, and well-being in elderly people.

SELF-ESTEEM has a well-recognized relationship topsychologica l wel l -be ing (Blascovich & Tomaka ,

1991). Self-esteem can be conceptualized as self-regard, anevaluat ion of one 's worthiness (Rosenberg, Schooler,Schoenbach, & Rosenberg, 1995). A number of scales havebeen designed to measure self-esteem (Blascovich &Tomaka, 1991), but there has been little research into thepsychometric properties of these scales, particularly usingdata from elderly people. The aim of this research was toassess the psychometric properties and.factor structure of aself-esteem scale suitable for the study of elderly adults,the Bachman revision (Bachman, 1970) of Rosenberg'sSelf-Esteem Scale (RSE; Rosenberg, 1965).

Rosenberg's Self-Esteem Scale (Rosenberg, 1965) is themost widely used measure of self-esteem (Blascovich &Tomaka, 1991). It was designed for research into adoles-cent self-esteem (Rosenberg, 1965). However, it has been

used in well-being research on all age groups, and is con-sidered to be an appropriate measure of self-esteem in theelderly (Breytspraak & George, 1979). Recent studies thathave used it with older adults include those of Duffy andMacDonald (1990) and Krause (1987).

Rosenberg (1965) designed the Self-Esteem Scale as aunidimensional measure of global self-esteem. Other re-searchers, building on his work, subsequently designedscales to measure different dimensions of self-esteem andin some cases domain-specific self-esteem (Blascovich &Tomaka, 1991). A number of principal components factoranalyses performed in the 1970s supported Rosenberg'sview that the RSE was unidimensional. However, some

investigators identified two highly correlated factors, posi-tive and negative self-esteem (reviews in Blascovich &Tomaka, 1991; Carmines & Zeller, 1979). In some studies,such as research into the psychological impact of unem-

ployment, positive self-esteem and negative self-esteemhave been used as separate variables (Feather & Bond,1983; Warr & Jackson, 1983). However, Carmines andZeller (1979) claimed that the putative two-factor structurewas likely to be a statistical artifact due to response set,since some items are positively worded while others arenegatively worded. They concluded that, instead of positiveand negative self-esteem, the two factors indicated by prin-cipal components analysis represented a global self-esteemfactor and systematic error variance due to the valence ofthe item wording. The negatively w orded items in the Self-

Esteem Scale seem to describe the opposite poles, on thesame continuum, to the positively worded ones rather thana conceptually different dimension. For instance, a posi-tively worded item ("I take a positive attitude towards my-self) is at the opposite pole to a negatively worded item("At times I think I am no good at all"), both items refer-ring to evaluation of worthiness. This could explain thehigh correlation between the two putative factors.

In the last few years there have been major developmentsin factor analysis. Confirmatory factor analysis (CFA), inwhich a theoretically derived factor structure is tested forits fit to the data, is now commonly used. Shahani, Dip-boy e, and Phillips (19 90) used confirmatory factor analysison the original RSE and reported that a structure with twofirst-order factors, self-derogation and self-enhancement(their terms for negative and positive self-esteem), was su-perior to a unidimensional one. Recently, Rosenbergseemed to have departed from his earlier unidimensionalviews concerning his scale and to have accepted the two-

factor model ascribed to by Shahani et al. (Rosenberg et al.,1995). A recent CFA study of the RSE was conducted byMarsh (1996), who used a modified seven-item version. Heargued that his results supported the reasoning of C armines

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USEFULNESS AND SELF-ESTEEM P97

and Zeller (1979), that the second putative factor was a sta-tistical artifact due to response set. Therefore, the evidenceis equivocal and the structure of the Self-Esteem Scale hasremained unclear.

The present study does not examine the original Self-Esteem Scale (Rosenberg, 1965) but the Bachman revision,a scale that arose from a large set of personality items usedin the Youth in Transition project (Bachman, 1970). Theitems of the revised scale are shown in Table 1. To date, theBachman revision has not been subjected to confirmatoryfactor analysis. The revision is similar to Rosenberg's Self-

Esteem Scale, both consisting of 10 items. However, the re-vision is an abbreviated composite of two measures, theSelf-Esteem Scale and a 10-item attitude scale devised byCobb, Brooks, Kasl, and Connelly (1966, in Bachman &O'Malley, 1977) for a study of adult workers who hadchanged jobs. In the process of reducing the composite

scale to 10 items, four items from the Self-Esteem Scaleand six from the scale of Cobb et al. were eliminated. TheCobb et al. items that were retained (items 7, 8, 9, and 10 inTable 1) were said to be "similar in content" to the Self-

Esteem Scale items that were replaced (Bachman & O'Mal-ley, 1977, p. 368). Bachman (1970) regarded the revisedscale as a unidimensional measure of global self-esteem.This view has been accepted in subsequent research thathas used the revision (Bachman & O 'Malley, 1986; Bynner,O'Malley, & Bachm an, 1 981; Rosenberg & Rosenberg,1978). However, the wording of the four replacement scaleitems emphasizes usefulness and competence, while thefour that were replaced (scale of Rosenberg, 1965) describe

positive self-regard. Hence, there is reason to suppose thefactor structure of the Bachman revision is different fromthat of the original RSE .

The B achman revision, (called RS E-B subsequently forthe sake of brevity) has been used extensively in research(e.g., Bachman & O'Malley, 1977, 1984, 1986; Bachman &Schulenberg, 19 93; Bynner et al., 1981; O'Malley & Bach-man, 1979, 1983; Rosenberg & Rosenberg, 1978; Rosen-berg et al., 1995; Wells & Sweeney, 1986). It has a reportedreliability of 0.75 (Cronbach's a) and good construct valid-ity (Bachman & O'Malley, 1977).

It was noted above that the four replacement items con-tain references to usefulness and competence. Specifically,items 8 and 9 in the R SE-B (see Table 1) contain words re-

Table 1. Bachman Revision (1970)of Rosenberg's Self-Esteem Scale

Item

1. I feel that I'm a person of worth, at least on an equal plane with others.

2. I feel that I have a numbe r of good qua lities.

3. I am able to do things as well as most other people.

4. I feel that I do not have much to be proud of.

5. I take a positive attitude towards myself.

6. I think I am no good at all.

*7 . I am a useful person to have around.

*8 . I feel I can't do anything right.

*9 . When I do a job, I do it well.*10. I feel that my life is not very useful.

Note: Scale items marked * were added by Bachman (1970).

lated to competence, whereas items 7 and 10 explicitlyrefer to usefulness. Of the six items retained from the RSE,item 3 seems to refer to competence and item 4 to achieve-ment, a concept related to usefulness and competence. The

references to usefulness could make the revised scale espe-cially applicable to the assessment of self-esteem in the el-derly, since to feel useful is thought to be important to thepsychological well-being of this group (Butler, 1985; Ryff,

1989). The four items not yet discussed (items 1, 2, 5, and6) refer to positive self-regard, the essence of self-esteem(Rosenberg et al., 1995). We therefore argue that the RSE-B may contain a usefulness/competence factor as well as aself-esteem factor.

To summarize the argum ent so far, two po ssible factor an-alytic models for the RSE-B could be derived from the pub-lished literature to date, which has assumed that the RSE-Brepresents the same construct as the RSE. The first possible

model comprises a unidimensional structure, measuringonly self-esteem (Bachman & O'Malley, 1986; Bynner etal., 1981; Rosenberg & Rosenberg, 1978). The secondmodel comprises two first-order factors, self-derogation andself-enhancement (following Shahani et al., 1990). A thirdmodel, derived from the history of the scale, in which an at-titude scale designed for adult workers was imported intothe existing self-esteem scale, and from an inspection of thewording of the items, comprises two first-order factors, posi-tive self-regard and usefulness/competence.

When performing confirmatory factor analysis, it is con-sidered good practice to develop a number of competingmodels and see which best fits the data (Hertzog, 1990;

Liang & Bollen, 1983; Marsh, 1996)," and this approachwas used in this research. The first two competing hypothe-ses were as follows.

Hypothesis I. The RSE-B is unidimensional and measures afactor called global self-esteem. The factor structure to testthis hypothesis was labeled model 1. All 10 items were spec-ified to load onto the global self-esteem factor.

Hypothesis 2. The RSE-B has a two-factor structure, the fac-tors being self-enhancement and self-derogation. The factorstructure to test this hypothesis was labeled model 2. Items1, 2, 3, 5, 1* and 9* were specified to load on the self-

enhancement factor, and items 4, 6, 8,* and 10* on the self-

derogation factor (* refers to items added by Bachman,1970).

Model 3 was derived from the history of the scale andfrom inspection of the wording of the scale items. The nextcompeting hypothesis was:

Hypothesis 3. The RSE-B has a two-factor structure, the fac-tors being positive self-regard and usefulness/competence.The first factor was labeled positive self-regard rather thanself-esteem to distinguish it from the general self-esteemfactor of models 4 and 5 (see below), but positive self-regardis conceptualized in this article as equivalent to self-esteem.Items 1, 2, 5, and 6 were specified to load on the positiveself-regard factor, and items 3, 4, 7,* 8,* 9,* and 10* on theusefulness/competence factor.

We also tested two m odels, each of which included a sec-ond-order factor. Second-order factors have been identifiedin addition to first-order factors in confirmatory factor anal-


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P98 RANZIJNETAL.

yses of a number of personality scales in recent years. Thesecond-order factor is usually considered to be superordi-nate to the first-order factors—in other words, to representa construct common to the first-order factors. For this rea-

son, models with second-order factors are often describedas hierarchical. For instance, Liang and Bollen (1983)found that Lawton's (1975) Philadelphia Geriatric CenterMorale Scale had a second-order "subjective well-being"factor, and McCallum, Mackinnon, Simons, and Simons(1995) found a scale factor "depression" in Radloff's(1977) Center for Epidemiological Studies-DepressionScale. However, a second-order factor could also be non-superordinate but simply the factor common to all the itemsof the scale. It is reasonable to state that any particular scaleis designed primarily to measure one major construct, andso all the items of the scale should be related to this con-struct. If this were not the case, researchers would use two

or more different scales rather than one. Therefore, it maybe the case that scales with at least two first-order factorscontain a general factor shared by all the items in additionto more specific factors each related only to some items.

The last two models tested were similar to models 2 and3, with the addition of the second-order factor. The hy-potheses therefore were:

Hypothesis 4. The RSE-B has a nested structure, with twofirst-order factors (self-derogation and self-enhancement)and one second-order factor (general self-esteem). The useof the term nested rather than hierarchical will be explainedlater. The factor structure to test Hypothesis 4 was labeledmodel 4. It differed from model 2 in that all 10 items were

specified to load on the general self-esteem factor in additionto their loading on the specific first-order factors.

Hypothesis 5. The RSE-B has a nested structure, with twofirst-order factors (positive self-regard and usefulness/com-petence) and one second-order factor (general self-esteem).The factor structure to test Hypothesis 5 was labeled model5. It differed from model 3 in that all items were also speci-fied to load onto the general self-esteem factor.

These models are structurally similar to one tested byGustaffson and Balke (1993), who suggested the termnested because the statistical procedure to test a nestedmodel involves the extraction of the first-order factors from

the residuals produced by extracting the second-order factorfrom the covariance matrix. However, whereas the first-order factors of Gustaffson and Balke's study were speci-fied to be orthogonal, in the present study they were mod-eled as correlated. This is because we theorized that thefirst-order factors would be related, possibly in a causalsense, to each other, with usefulness/comp etence influenc-ing positive self-regard. A recent article by Mutran, Reitzes,Bratton, and Fernandez (1997) concluded that competencehas a strong influence on self-esteem, which provides sup-port for such a causal relationship among the factors in theRSE-B. In spite of the difference between our model speci-fication and that of Gustaffson and Balke (1993), we feel

that the term nested is appropriate for our models becausethe analytical procedure used is basically the same.

In similar analyses previously reported, the term hierar-

chical has sometimes been used (e.g., Liang & Bollen,

1983; McCallum et al., 1995). In a true hierarchical struc-ture the first-order factors are correlated with the second-order factor and the individual items do not load directlyonto the second-order factor. A hierarchical structure was

rejected in this case for two reasons. First, the second-orderfactor, general self-esteem, could not be superordinate topositive self-regard because these two constructs are con-ceptually equivalent. Second, we thought that general self-

esteem was common to all the scale items, and thereforeeach item was specified to load onto the second-order factoras well as one of the first-order factors.

In summary, five competing m odels w ere constructed forthe Bachman revision (1970) of the Self-Esteem Scale(Rosenberg, 1965). Two of the models contained a factor,usefulness/competence, that had not previously been asso-ciated with this scale. The first three models containedsingle-order factors and the final two were nested models

that also contained a second-order factor. The data were de-rived from a longitudinal study of ageing, and the modelswere validated with data collected 2 years after the firstwave of the survey. It was expected that model 5 wouldprovide the best fit to the data.

METHOD

Participants and Procedure

Data for this study were provided by 1,087 elderly peo-ple (mean age 77.43, range 70-103 years) who comprised asubsample of the Australian Longitudinal Study of Ageing(ALSA; Centre for Ageing Studies, 1992). The ALSA is a

multidimensional, multidisciplinary research project thataims to increase understanding of how social, biomedical,behavioral, economic, and environmental factors are relatedto age-related changes in the health and well-being of el-derly Australians. From late 1992 to early 1993, compre-hensive baseline interviews, covering a broad range of de-mographic, health, social, and psychological issues, wereconducted with 2,087 elderly adults living in the Adelaidemetropolitan area. The ALSA also included a series of self-

complete questionnaires, filled in by the participants afterthe interviews, which included the self-esteem scale. Longi-tudinal follow-up occurred 2 years later. Details of theALSA have been published elsewhere (Centre for Ageing

Studies, 1992; Clark & Bond , 1995; Mawby, Clark, Kalucy,Hobbin, & Andrews, 1996; Ranzijn & Luszcz, 1994).

The target sample for the ALSA (N = 3,623) was ran-domly generated from the South Australian Electoral Rolland was stratified by sex and 5-year age cohorts from 70 upto the age of 85 years and over. Older males were oversam-pled, as were the oldest groups, in an attempt to obtain rea-sonable numbers of each of these in subsequent waves ofthe study. It could be argued that the stratified nature of thesampling plan should have been taken into account so as tominimize biases in subsequent analyses. However, it hasbeen found by Mawby et al. (1996), who used the samesample of 1,799 peop le, that there were no significant dif-

ferences in incom e, AD LS, IADL S, hospitalizations or self-rated health between analyses using raw data and thoseweighted to adjust for stratification. To test for possible bi-ases in self-esteem scores, both weighted and unweighted


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data were used to analyze the differences in the meanscores for the four age groups and for men and women (seeResults). The values of mean scores for each cell were vir-tually identical. However, the very small differences be-

tween age groups became significant because of the verylarge number of cases estimated through the weighting pro-cedure. Therefore, unweighted data were used for the factoranalyses and subsequent analyses.

Of the 3,263 people approached for the first wave of theALSA, 558 were ineligible for reasons such as death, incor-rect age or address, being outside the geographic region,and language difficulties. Of the 2,705 in the eligible sam-ple, 1,477 (54.6%) participated. Of the nonrespondents, 780(28.8%) refused outright, the 2-hour length of the interviewand the long-term involvement sought by the ALSA design-ers being the major deterrents to participation. The othermajor reason for nonparticipation was illness of self or

spouse (16.2%). Spouses older than 65 and other householdresidents over the age of 70 were also invited to participate,resulting in another 610 participants and hence a totalALSA sample of 2,087. This number included people liv-ing in institutional care as well as in the community. How-ever, we were interested only in community-dwelling peo-ple aged 70 years or more (N = 1,799). Full details of thesampling frame and response statistics have been publishedby the Centre for Ageing Studies (1992). Of the first-wavepool of 2,087 people, 241 were deceased and 42 had movedor could not be contacted at the 2-year follow-up, leaving1,804 people. One hundred twenty-five refused to partici-pate again, resulting in a retention rate of 93% (N = 1,679).

The self-esteem data used in this report were obtainedfrom the optional self-complete questionnaire. Consequently,fewer people answered items from this questionnaire (N =1,297 at the first wave and 957 at the 2-year follow-up) com-pared to those who took part in the interview (N = 1,799 and1,331, respectively). There were 1,087 people (84% of thoseattempting the self-complete questionnaires) who fully com-pleted all the self-esteem items at the baseline survey of theALSA and 875 at the follow-up (91%), and they comprisedthe subsamples for the analyses reported here.

Descriptive statistics from the baseline data collection ondemographic data and health and function for the full sam-ple, for the sample that completed the full self-esteemscale, and for those who completed only some optionalquestionnaire items are shown in Table 2. In addition to thevariables shown, there were no differences in the propor-tions of male respondents, the percentages being 54.2%,54.9%, and 5 0.5% respectively.

Table 2 shows there were few differences between thegroups. The full self-esteem group was younger, had higherincome, more education, fewer ADLs and IADLs, betterself-rated health, less depression, and better cognitive func-tion than the other two groups. The differences between thefull sample and full self-esteem samples were less thanthose between the full self-esteem and partial optionalquestionnaire samples. However, the differences in everycase were small.

Materials

The Bachman revision (1970) of the Self-Esteem Scale

Table 2. Baseline Descriptive Statisticsfor the Full Sample and Self-Esteem Subsamples

Variable

Age (years)

Income

Education

Hospital

ADLs

IADLs

S/R Health

Depression

Minimental

Full Sample

N

1,799

,686

,783

1,799

,791

,790

,795

,717

,767

M(S.D.)

78.35 (5.89)

2.96(1.04)

3.70(1.38)

0.38 (0.90)

0.34 (0.93)

0.89(1.68)

2.92(1.11)

7.95 (7.09)

18.89(2.78)

FullSelf-Esteem

N

,087

,044

,087

1,087

,087

,086

,086

,059

,079

M(S.D.)

77.43 (5.52)

3.05(1.05)

3.82(1.37)

0.34 (0.79)

0.28 (0.84)

0.74(1.46)

2.83(1.10)

7.48 (6.82)

19.33(2.18)

SomeSelf-Esteem

N

21 0

195

21 0

21 0

21 0

21 0

21 0

20 0

20 6

M (S.D.)

80.81 (6.26)

2.71 (0.87)

3.48(1.29)

0.48(1.21)

0.48(1.15)

1.14(1.83)

3.06(1.09)

8.43 (7.40)

18.36(2.70)

Note: For income, a score of 2 represents an annual income range of$5,001-$12,000, whereas 3 is equivalent to $12,001-$20,000 Australiandollars. For education, a score of 3 means the participant left school at 14

years, a score of 4 means 15 years. Hospital = number of times in hospitalin the previous 12 months. ADLs and IADLs are measures of everydayfunction: ADLs = number of Activities of Daily Living problems, IADLs= number of Instrumental Activities of Daily Living problems (Duke Uni-versity, 1978). Self-rated health is the score on a single-item rating rang-ing from 1 (excellent) to 5 (poor). Depression = score on C ES -D(Radloff, 1977); a score of 8 indicates low depression. Minimental =score on items from the Mini-Mental State Exam (Folstein et al., 1985); ascore of 18 indicates good cognitive function.

(Rosenberg, 1965) used in this research comprised the 10items listed in Table 1. Participants were asked to reporthow often each statement was true for them. The responsecategories were "almo st always true " (1), "often tru e" (2),

"sometimes tru e" (3), "not often true" (4), and "never true"(5). Items 1, 2, 3, 5, 7, and 9 were reverse-scored in the pro -cess of data entry. Scores were summed and the total scoredivided by 10, and subscale scores divided by their numberof items, to give a possible rang e from 1 to 5. Higher scoresindicate higher self-esteem. A reliability analysis of the fullscale showed that the RSE-B had good reliability, Cron-bach's a = 80.

The following variables were assessed to confirm the in-dependence of the first-order factors, using the method ofCarmines and Zeller (1979). Depression was measured bythe Center for Epidemiological Studies-Depression Scale(Radloff, 1977), morale was measured by 15 items from thePhiladelphia Geriatric Center Morale Scale (Lawton, 1975),and perceived control by items from the Expectancy ofControl subscale of the Desired Control Measure (Reid &Zeigler, 1981). Further details of the adapted scales used inthe ALSA can be found in Luszcz (1996). It was found dur-ing data analysis that depression and morale had skeweddistributions, so transformed scores were used in subse-quent analyses. For depression, the scores underwent a log-arithmic transformation, and for morale reflected andsquare rooted scores were used. Finally, two subscales ofthe Adelaide Activities Profile (Clark & Bond, 1995),namely, the Domestic Chores and Home Maintenance sub-scales, were used to measure functional ability to partici-pate in everyday activities.

The participants completed the interview and other scalesin their own homes. The initial data preparation was per-formed with SPSS-X (SPSS/Norusis, 1993) and the confir-


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P100 RANZIJNETAL.

matory factor analysis using LISREL for Windows (Joreskog& Sorbom, 1993).

Analysis

PRELIS2, the preprocessor for LISREL8 (Joreskog &Sorbom, 1993), was used to generate the matrices for sub-sequent analysis by LISREL8. All RSE-B items were de-clared to be ordinal, because there were only a small num-ber of response categories and the intervals betweenadjacent categories could not be assumed to be equal. Ac-cordingly, the correlation matrix generated consisted ofpolychoric correlations (Joreskog & Sorbom, 1993). Theevaluation of goodness of fit was performed by LISREL8with the generally weighted least-squares method, using theasymptotic covariance matrix.

RESULTS

The mean total self-esteem score was 4.14 (SD .55). In-dividual item means ranged from 3.74 (item 4) to 4.50(item 6), with standard deviations between 0.73 (item 2)and 1.20 (item 4). There were statistically significant differ-ences between different age groups, F(3,1083) = 4.73, p <

.01. Post-hoc Scheffe analyses showed that people aged 85years or more had a lower mean self-esteem score (4.01,SD .57) than the 70-74 (M 4.19, SD .52) and 75-79 yearage groups (M 4.16, SD .56), but the differences were verysmall. There were no significant differences in the meanscores of men compared to women. As mentioned earlier,analyses using weighted data produced results virtuallyidentical to those using unweighted data (the equivalent

mean scores to those mentioned above were 4.01 , 4.19, and4.18) so unweighted data have been used for all subsequentanalyses.

The polychoric correlation m atrix is shown in Table 3.

Confirmatory Factor Analyses

In CFA, the assessment of the fit of a model is based onthe relative importance attached to the various indices pro-duced by the program used, a debate that is at present un re-solved (for some points of view, see Bentler, 1990; Bollen,1990; Joreskog & Sorbom, 1993; McDonald & Marsh,1990; Marsh, 1996; Wade et al., 1996). Some authoritiesconsider that the best indicators are the root-mean-square

error of approximation (RMSEA), which should be below.05 and as low as possible, and the probability that theRMSEA is less than .05, this probability being as close as

Table 3. Polychoric Correlation M atrix of RSE-B

Item Item Item Item Item Item Item Item Item

1 2 3 4 5 6 7 8 9

Item 2 .74

Item 3 .54 .51

Item 4 .25 .23 .19

Item 5 .49 .51 .45 .25

Item 6 .33 .30 .21 .48 .38

Item 7 .45 .56 .52 .23 .46 .31Item 8 .26 .24 .31 .37 .32 .56 .30

Item 9 .37 .37 .48 .21 .41 .30 .56 .32

Item 10 .29 .27 .29 .42 .40 .55 .35 .54 .29

possible to 1.00 (Gustaffson & Stahl, 1996). A lowerRMSEA and a higher p value indicate that one model is su-perior to another. One model is also considered better thananother if its x2 value is lower, the probability of this x2

value is higher, the x2 df ratio is lower, the root-mean-square residuals index (RMR) is lower, and the goodness-of-fit indices are higher, preferably close to 1.00 (Joreskog& Sorbom , 1993), than the values of the other model. How-ever, the final determination of best fit should be theoreti-cally based as well as statistically driven (Bollen, 1990;Breckler, 1990; Browne & Cudeck, 1989; Hertzog, 1990).In the analyses reported here, no modifications were per-formed on the models in the interests of achieving a betterfit. This decision was taken because it was thought moreimportant to maintain theoretical consistency than to usethe results of model testing to adjust the models in order toimprove the fit, the latter being a statistically driven rather

than theoretically driven approach. The results of the con-firmatory factor analyses are summarized in Table 4.

Table 4 shows that the only models that are well fittingaccording to the RMSEA criteria (RMSEA < .05, and p ofRMSEA > .05) are models 4 and 5. Model 2 (the first two-factor model) is superior to model 1 (the unidim ensionalmodel) in every respect except PGFI (parsimony goodnessof fit), where the results are the same. Model 2 is also supe-rior to model 3 (the competing two-factor model). Model 4(the first nested model) is superior to model 2 in every re-spect except PGFI, and the fit is excellent according to allthe criteria. The PGFI in m odel 4 is lower than in mo dels 2and 3 because of its increased complexity, having a second-

order factor, but is still acceptable. The reduction in x

2

formodel 4 compared to model 2 is significant, x2(l) = 8.77, p< .001 (two-tailed). Model 5 has a higher x2 value, df ratio,and RMSEA than model 4, and is equal on the other crite-ria. The reduction in x2 from model 3 to model 5 is signifi-cant , X2(l) = 16.04, p = < .001 (tw o-tailed).

Using these indices alone, model 4 seems superior tomodel 5. However, some of the factor loadings for model 4were not meaningful, whereas all loadings for model 5were meaningful (see Table 5). The determination of thefit of a model should, in addition to the other criteria,be based on the meaningfulness of the factor loadings

Table 4. Results of ModelTesting of RSE-B

Model x2 df x2/df RMSEA (p) RMR GFI AGFI PGFI

Testing with Wave 1 ALSA data

1 274.63 35 7.85 .079 (~0) .14 .96 .94 .61

2 143.41 34 4.22 .054 (.20) .07 .98 .97 .61

3 230.82 34 6.79 .073 (~0) .13 .97 .95 .60

4 55.72 24 2.32 .035 (.98) .037 .99 .98 .43

5 70.38 24 2.93 .042 (.86) .037 .99 .98 .43

Validation with Wave 3 ALSA data

4 66.22 24 2.76 .045 (.73) .050 .99 .98 .43

5 63.01 24 2.62 .043 (.79) .051 .99 .98 .43

Note: RMSEA = root-mean-square error of approximation; RMR =root-mean-square of residuals; GFI = Goodness-of-Fit Index; AGFI = Ad-justed Goodness of Fit (goodness of fit adjusted to take into account thedegrees of freedom); PGFI = Parsimony Goodness of Fit (goodness of fitadjusted to take into account the complexity of the model).


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(Joreskog & Sorbom, 1993). On balance, model 5 wasthought to be the best-fitting mod el. In most cases the first-order loadings were high, and in all cases they were sig-nificant. The sum of the first-order factor variances ex-

ceeded the variance of the second-order factor. In otherwords, the contribution of the first-order factors to the fac-tor structure was greater than that of the second-order fac-tor. This means that the first-order factors are clearly iden-tified by the analysis, given that the second-order factorhas already been extracted. The phi coefficient (the corre-lation between the first-order factors) for model 5 was .84(p < .001).

Cross- Validation

The factor structures of models 4 and 5, the models thatprovided the best fits to the baseline ALSA data, were vali-dated with the follow-up data obtained 2 years later. The re-

sults of cross-validation are shown in Table 4. Model 5 wassuperior on all the criteria except RMR , where m odel 4 wasmarginally superior. A number of the factor loadings formodel 4 were again unstable, whereas all the loadings formodel 5 were meaningful (see Table 5). It was concludedfrom the analyses of both data sets that model 5 providedthe best fit to the data.

Table 5. Factor Loadings for M odels 4 and 5: \ Weights

Model 4

First-Wave Subsam ple Third-Wave Subsam ple

Item S E S D G S E S E S D G S E

.84***

- . 03

.02

.87***

.86***

.30

.18

.62*** .35* .73*** - .14

.57*** .14 .63*** - .0159*** 34* 74*** _ 0 8

.73*** .25* .%!*** .08

.54*** .53*** .72*** - .36

.63*** .36*** .72*** .02

.45*** .56*** .69*** - .39

10 .65*** .35*** .75*** - .11

Variance (I \ 2 ) 2.71 1.68 1.17 3.57 2.23 .45

Model 5

F irs t-W av e S ub sa mp le T hird -W av e S ub sa mp le

Item PSR U/C GS E PSR U/C GSE

2

3

4

5

6

7

8

9

10

. 8 5 * * *

. 8 5 * * *

.62***

.30*

.76***

.23*

.71

.29**

.63***

.33**

.12

.09

.08

.55***

.29**74***

.20

.65***

.22

.66***

.85***

.84***

.66***

.26*

.74***

.19*

.76***

.21 *

.77***

.26**

.27*

.26*

.23*

59***37***

.86*

.23*

.71***

.14

7 1 * * *

Variance (IX 2) 1.91 1.62 1.91 1.93 1.86 2 .49

Note: SE = self-enhancement, SD = self-derogation, PSR = positiveself-regard, U/C = usefulness/competence, GSE = general self-esteem.

*p < .05; **/>< .01; ***/>< .00 1.

Analyses Using Subscales

The positive self-regard subscale was computed by usingscores on items 1, 2, 5, and 6, and the usefulness/compe-tence subscale consisted of items 3, 4, 7, 8, 9, and 10.

Higher scores indicate more positive self-regard and astronger sense of usefulness/competence, respectively. Sub-scale scores were divided by their number of items, to givea possible range from 1 to 5. This was done to enable com -parisons to be made with the full scale scores. The reliabil-ity of the positive self-regard subscale was 0.67, and that ofthe usefulness/competence subscale was 0.69 (Cronbach'sa) , values that are quite adequate for scales with only a fewitems (Nunnally, 1978). The subscale score for positiveself-regard was 4.34 (SD 0.57), and that for usefulness/competence was 4.09 (0.59), mean scores being high inboth cases (range 1-5).

As a final test of the construct validity of the first-order

factors, the subscales were correlated with three variablesthat have been associated with self-esteem, namely depres-sion, morale, and personal control (Andrews & Robinson,1991), and also with functional ability. The resulting corre-lations of the full self-esteem scale with these variableswere all significant, p < .001 (see Table 6).

If two factors are truly distinct, it would be expected thatthey would be differentially related to relevant variables(Carmines & Zeller, 1979). Following the procedure usedby Carmines and Zeller in their factor analysis of the Self-

Esteem Scale (Rosenberg, 1965), correlations were com-puted between each subscale and the variables mentionedabove. These correlations are also shown in Table 6. Analy-

ses of the differences between the correlations, using themethod for determining the significance of the differencebetween dependent correlations described by Cohen andCohen (1983), showed there were no differences on depres-sion and control for the first-wave subsample or on do mes-tic chores for the third-wave subsample, but there were dif-ferences between all the other correlations. Because someof the correlations were noticeably larger in the third-wave

Table 6. Matrix of Pearson Correlationsof Full Self-Esteem Scale and Subscales with Weil-Being,

Control, and Functional Ability Variables

Variable

Self-esteem

Positive self-regard

Usefulness/competence

Difference

Self-esteem

Positive self-regard

Usefulness/competence

Difference

First-Wave Subsample

Dep

- .25

- .20

- .25

ns

M or

.39

.30

.38

Cont

.53

.46

.48

ns

Third Wave Subsample

- .46

- .37

- .44

.46

.34

.46

.59

.50

.56

p < . 0 5

DC

.14

.09**

.15

p < .001

.12

.08*

16

p < .001

HM

.20

.11

.22

p < .001

.23

.16

.25

p<.00\

Note: Dep = depression, Mor = morale, Cont = personal control, DC =

domestic chores subscale, and HM = household maintenance subscale ofAdelaide Activities Profile; Difference = difference between correlationswith subscales.

*p < .05; ** p < .01; all other correlations are significant, /? < .001 .


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P102 RANZIJNETAL.

subsample than in the first-wave subsample, the first-wavecorrelations were repeated using only those people w ho had

full self-esteem data at both waves of the study (N = 875).

However, the results were virtually identical, the values of

the correlations differing only between 0 and .02.

DISCUSSION

This study has shown that the Bachman revision (1970)of the Rosenberg Self-Esteem Scale (1965) (RSE-B) is a

multidimensional scale with two first-order factors, labeledpositive self-regard and usefulness/competence, and an or-

thogonal second-order factor, general self-esteem.The original Self-Esteem Scale of Rosenberg (1965) was

conceptualized as unidimensional, and this has been verifiedrecently by Marsh (1996). However, the Bachman revision(1970) is a different scale, with four items not in the originalscale, all construed as referring to self-perceived usefulness

or competence. In this study, the unidimensional model(model 1) wa s the worst fitting of all five models tested.

The model based on the division of the scale into self-

derogation and self-enhancement (model 2) provided a bet-

ter fit to the data than the unidimensional model, but not as

good as later models. If one model demonstrates a better fit

than another, the superior m odel is the only one that can be

accepted as valid. We therefore agree with Carmines and

Zeller (1979) and Marsh (1996) that self-derogation and

self-enhancement are not separate factors.Th e fit of the third model, with two first-order factors,

positive self-regard and usefulness/competence, was aboutthe same as model 1 but not as good as either model 2, 4, or

5, and was therefore rejected. The fourth model tested, anested model with a second-order general self-esteem factorin addition to the first-order factors self-derogation and self-

enhancement, provided a better fit than the previous threemodels, but not as good as model 5, so it too was rejected.Furthermore, model 4 was also unacceptable because the

factor loadings were not meaningful. Cross-validation veri-fied that model 5 was superior to model 4.

Model 5 was a nested model, with positive self-regardand usefulness/competence as first-order factors and a sec-

ond-order factor, general self-esteem. The finding that a

nested model with a second-order factor provides the bestfit is consistent with previous studies of other scales in

which a general factor has been found (e.g., Liang &Bollen, 1983; McCallum et al., 1995). We conclude that the

RSE-B contains a factor common to all items as well as

their specific first-order factors. It could generally be ar-

gued that all scales that are not unidimensional should con-

tain such a common core, and therefore a nested model is

the most app ropriate one to test with CFA.

The people in this study had high scores on both the pos-

itive self-regard and usefulness/competence subscales. It

seems that perceived usefulness is not necessarily a func-tion of economic productivity, since the people in this studyhad been retired for an average of 12 years. It may be mis-

guided to link usefulness with economic productivity when

discussing elderly people. On the other hand, the respon-dents may have adjusted psychologically to any negative ef-

fects of retirement (see Lazarus & DeLongis, 1983, for a

discussion of adjustment mechanisms in the elderly). These

high scores, however, may not accurately reflect the widerpopulation of elderly Adelaide people, given that the re-

sponse rate was not as high as would have been wished.Nevertheless, outright refusals comprised just over a quar-

ter of those people initially approached, and this was likelyto have been due to the long-term time commitment re-

quested and the length of the interview and other aspects of

the questionnaire. Th erefore, we concluded that the data are

a reasonable reflection of the population from which the

sample was drawn.The usefulness/competence factor has not previously

been identified in studies of self-esteem. However, the con-

cepts of usefulness, competence, achievement, and produc-tive activity have lately received increasing attention in the

gerontological literature. For example, Stevens (1993)found that involvement with others was significantly relatedto feeling useful. Mclntosh and Danigelis (1995) found an

association between informal volunteering (such as per-forming chores for friends) and well-being in adults overthe age of 60 years. Furthermore, Mutran et al. (1997)found a strong association between competence, measuredby a single item of self-assessment, and self-esteem. Re-

search on younger adults has also shown a relationship of

self-esteem with competence (Feather, 1991a, 1991b) and

sense of structure and purpose in the use of time (Bond &

Feather, 1988; Feather & Bond, 1983), which may be re-

lated to usefulness. The relationships among usefulness,competence, purpose, and self-esteem in older people de-

serves further exploration. The possible utility of the use-

fulness/competence subscale for future research is indicated

by its reliability, which is quite high for a scale with so fewitems. The scale could be further developed by the additionof other relevant items reflecting different aspects of useful-ness, competence, and achievement.

As a further test of the robustness of the factor structure,the differences in the correlations of the factors with othervariables relevant to self-esteem were tested for signifi-cance. These variables were depression, morale, personalcontrol, and functional ability. Functional ability was as-

sessed by two subscales, measuring the extent to whichpeople carried out normal domestic chores and home main-tenance. The findings indicated that the factors were con-

ceptually distinct. Usefulness/competence had a stronger

relationship to the indicator variables than did positive self-regard. The possible causal pathways here also deserve fur-

ther investigation. Usefulness/competence may be a goodpredictor of both well-being and personal control. Morework could also be done on the association between useful-ness/competence and functional ability. There could be a

causal pathway from functional ability to well-being via

usefulness/competence and/or positive self-regard.

One puzzling finding was that some of the correlations re-

ferred to in the previous paragraph were substantially higherin the third-wave subsample than in the first wave. One pos-

sible explanation was that the subsamples were not the same,since the first wave contained over 200 people who were not

in the third wave. However, even when the correlations wererecalculated for only those first wave respondents who werealso in the third wave, the results were virtually identical to

those using the full subsample. Two details should be noted.


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The trends are the same for both waves, and the relationshipswith functional activity are more stable than those w ith well-being and control. Perhaps the strengthened correlations hadsomething to do with the respondents being 2 years older and

more familiar w ith the inventories. It is not clear why the fac-tors appear to become more differentiated. Overall, the vali-dation analyses strengthen support for the conceptual dis-tinctness of the two factors.

In summary, this study has demonstrated that the Bach-man revision (1970) of the Self-Esteem Scale (Rosenberg,1965) has a nested structure, with a second-order scale fac-tor in addition to two first-order factors, usefulness/compe-tence and positive self-regard. The newly identified useful-ness/competence subscale, a six-item scale with adequatereliability, could be further developed and gainfully em-ployed in future research on elderly people. Little work hasbeen done so far into the importance of feeling useful and

competent to the well-being of elderly people. This studysupports the view that feeling useful and competent may berelated in important ways to the self-esteem of older adults(Mcintosh & Danigelis, 1995; Mutran et al., 1997; Ryff,

1989). Preliminary results also indicate that usefulness/competence may be an important predictor of well-being.Further research is required to clarify the relationshipsamong usefulness, competence, and self-esteem in elderlypeople and the ways in which they affect well-being.

ACKNOWLEDGMENTS

, The authors gratefully acknowledge the work of the Centre for Ageing

Studies, Flinders University of South Australia, which, under the directionof Prof. Gary Andrews, carried out the Australian Longitudinal Study ofAgeing (ALSA) that was the source of the data for this study. This re-search was funded in part by a grant from the U.S. National Institutes ofHealth (Grant AG 08523-02) and by grants from the Sandoz Foundationfor Gerontological Research, the South Australian Health Commission,and the Australian Rotary H ealth Research Fund. We are also grateful toDrs. J. G. Bachman and P. M. O'Malley and two anonymous reviewers fortheir helpful and detailed comments.

Address correspondence to Dr. Rob Ranzijn, School of Psychology, TheFlinders University of South Australia, GPO Box 210 0, Adelaide 500 1,Australia. E-mail: [email protected]

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Received July 19, 1996

Accepted September 19, 1997

Nominations for Prestigious Awards

DEADLINE...MAY 8,1998Robert W. KleemeierAward

Donald P. KentAward

Glenn FoundationAward

Nathan Shock NewInvestigator A ward

PGC Polisher R esearchInstitute Award

To a Fellow of The Gerontological Society of America in

recognition of outstanding research in the field of gerontology.

To a Fellow of The Gerontological Society of America who best

exem plifies the highest standards for professiona l leadership in

gerontology through teaching, service, and interpretation of

gerontology to the larger society.

Open to all scientists, regardless of field or nationality, for

significant research contributions to the biology of aging.

For outstanding contributions to new know ledge about aging

through basic biological research.

Open to all disciplines, for a significant contribution from

applied research that has benefited older people and their care.

For further information about these and additional awards, contact:

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