Scandinavian Journal of Psychology, 2000, 41, 307± 314
Development and validation of Swedish classical and modern
sexism scales
BO EKEHAMMAR, NAZAR AKRAMI and TADESSE ARAYA
Department of Psychology, Uppsala University, Sweden
Ekehammar, B., Akrami, N. & Araya, T. (2000). Development and validation of Swedish classical and modern sexism scales. Scandinavian
Journal of Psychology, 41, 307± 314.
In two studies we develop and validate a ClassicalÐovert or directÐand a ModernÐ covert or subtleÐSexism Scale concerning attitudes
toward women, for a Swedish (Scandinavian) context. Further, we examine whether these two forms of prejudice are distinguishable.
Confirmatory factor analyses showed that, although highly correlated, classical and modern sexism are distinguishable. The construct
validations showed that men had higher means on modern and classical sexism scores than women, and that our scales were related to other
constructs as expected. In a third study, we analyzed the knowledge and the content of cultural stereotypes about women. There were no
differences in the knowledge of cultural stereotypes between men and women or between high- and low-sexist individuals. The findings are
discussed in relation to previous international studies that examine people's modern and=or classical sexism.
Key words: Modern sexism, classical sexism, construct validity, gender differences, prejudice.
Bo Ekehammar, Department of Psychology, Uppsala University, Box 1225, SE-751 42 Uppsala, Sweden. E-mail: [email protected]
INTRODUCTION
In the last few decades, much has been written about
stereotyping and prejudice. However, most of the research
has mainly focused on racism directed toward African
Americans (e.g., Franco & Maass, 1999; Pettigrew &
Meertens, 1995; Sears, 1988; Swim, Aikin, Hall & Hunter,
1995). Recently, several researchers have begun to pay
attention to another type of prevalent prejudiceÐ sexism
(e.g., Benokraitis & Feagin, 1995; Campbell, Schellenberg &
Senn, 1997; Eagly & Mladinic, 1989; Frable, 1989; Glick &
Fiske, 1996; Reid, 1988; Swim et al., 1995; for a review, see
Deaux & LaFrance, 1998). Sexism is generally understood
as endorsement of prejudicial beliefs based on a person's
gender (cf. Campbell et al., 1997; Frable, 1989). However, in
the present study, and following Glick and Fiske (1996),
Reid (1988), and Swim et al. (1995), we define sexism as
prejudice or discrimination directed against women.
Sexist attitudes are part of the history of our civilization,
where women are often discriminated against to varying
degrees in different societies, for example in the political
arena (Reid, 1988), in the employment market, and in salary
levels. Further, family roles are most often unequally
divided even for women with professional careers (Biernat
& Wortman, 1991). In most countries, women are also
underrepresented (if represented at all) in leading positions
in both private and public sectors (e.g., in Sweden, see
Statistics Sweden, 1999).
Drawing insight from the research on racial prejudice,
some researchers have argued that sexism, like racism, is
manifested in different forms (e.g., Glick & Fiske, 1996;
Swim & Cohen, 1997; Swim et al., 1995; Tougas, Brown,
Beaton & Joly, 1995). Changes in sociopolitical climate
(Katz & Hass, 1988; Van Ommeren & Ishiyama, 1998) have
probably led to the abandonment of traditionally endorsed
values. Additionally, people's tendency to present themselves as non-prejudiced and socially or politically ``correct''
(Crosby, Bromley & Saxe, 1980) may prevent the expression
of blatant forms of prejudice (Pettigrew & Meertens, 1995;
Swim et al., 1995). Moreover, in many countries, direct
expression of racial and sexist prejudice may lead to legal
punishment, and therefore some people can be reluctant to
express their prejudice openly (Akrami, Ekehammar &
Araya, 2000; Franco & Maass, 1999). Accordingly, some
researchers argue that blatantly negative attitudes toward
women have decreased over the past few decades (Campbell
et al., 1997; Mason & Lu, 1993). However, there are reasons
to suspect that the covert expression of these attitudes has
not decreased (Swim et al., 1995; Tougas et al., 1995).
Thus, to tap these recent changes in prejudicial attitudes,
researchers have begun to make a distinction between the
old-fashioned and modern forms of prejudice. For example,
Swim et al. (1995) distinguish old-fashioned from modern
sexism, echoing the distinction made in racism (e.g.,
McConahay, 1986; Sears, 1988). Similarly, Tougas et al.
(1995) introduced the concept neosexism, akin to ambivalent
racism (Katz & Hass, 1988), and defined it as a ``manifestation of a conflict between egalitarian values and residual
negative feelings toward women'' (p. 843). Further, Glick
and Fiske (1996) consider sexism as a multidimensional
construct consisting of two ambivalent attitudesÐ benevolent and hostile Ðwhere benevolent sexism may be used to
indirectly compensate the hostile form. In the present study,
we use the terms modern and classical to denote the two
forms of sexist prejudice.
Although women are not a minority group (Hacker,
1981), they are historically considered as a low-power group
# 2000 The Scandinavian Psychological Associations. Published by Blackwell Publishers, 108 Cowley Road, Oxford OX4 1JF, UK and
350 Main Street, Malden, MA 02148, USA. ISSN 0036-5564.
308
B. Ekehammar, N. Akrami and T. Araya
and are often victims of discrimination (e.g., Glick & Fiske,
1996). Thus, sexism has been frequently associated and
compared with other forms of prejudice, for example, racism
(e.g., Dovidio & Gaertner, 1998; Reid, 1988). Therefore,
some researchers emphasize the similarities between racism
and sexism both in the specific underlying beliefs and forms
of expression. We argue, following Sears (1988) and Swim et
al. (1995), that the underlying elements of modern racism
are also valid for modern sexism. These include (a) denial of
continued discrimination, (b) antagonism toward outgroups' demands, and (c) lack of support for policies
designed to help outgroups.
Recently, some researchers studying the automatic and
controlled components of racism have emphasized the
necessity of exploring the differences between high- and
low-prejudiced people in their knowledge of the contents of
cultural stereotypes (e.g., Augoustinos, Ahrens & Innes,
1994; Devine, 1989; Lepore & Brown, 1997). Devine (1989,
Study 1), for example, asked participants to list the cultural
stereotypes they knew about African Americans. Listing
mostly negative stereotypes, high- and low-prejudiced
individuals did not differ in their knowledge about the
cultural stereotypes against African Americans. These
studies suggest the need of similar exploration of the content
of cultural stereotypes about women. To the best of our
knowledge, no research has undertaken a close look at this
issue.
The discussion above underscores the importance of
developing instruments that tap the covert forms of sexism.
Therefore, the primary aim of the present study is to develop
a classical and a modern sexism scale for a Swedish
(Scandinavian) context and examine the construct validity
of these scales. Further, we examine the content of the
cultural stereotypes.
STUDY 1: CONSTRUCTING CLASSICAL AND
MODERN SEXISM SCALES
We devised a pool of items chosen to reflect Swedish beliefs
and ideas underlying classical and modern sexism (see
method section). We compare modern and classical sexism,
and we expect responses to the two sets of items to be
correlated. In addition, we anticipate that the responses will
be characterized by a two-factor structure, one factor
representing modern sexism and the other classical sexism.
Extant research indicates that racism and sexism are
connected (e.g., Dovidio et al., 1998; Sidanius, 1993) and
that men have higher sexism scores than women (Swim et
al., 1995). In fact, most tests of construct validity of sexism
scales have relied on determining whether men and women
respond differently to the scales (DelBoca, Ashmore &
McManus, 1986; Swim et al., 1995). Further, based on the
findings of previous research, we expect the sexism scales to
be positively correlated with the racism and the conservatism scales (e.g., Ficarrotto, 1990; Glick & Fiske, 1996;
# 2000 The Scandinavian Psychological Associations.
Scand J Psychol 41 (2000)
Pratto, Sidanius, Stallworth & Malle, 1994; Sidanius, 1993).
For example, Fazio, Jackson, Dunton, and Williams (1995)
have argued that conservatism is a significant predictor of
racism. Moreover, following Pratto et al. (1994), we consider
conservatism as a legitimizing myth that separates people
into different groups and accords a positive social value for
the preferred but a negative value for the non-preferred
group. Women, we suggest, are one example of such a nonpreferred group.
Consequently, in our construct validation of the scales we
expect a) that men display higher sexism scores than women,
and b) a positive correlation between modern and classical
sexism on the one hand and conservatism, and modern and
classical racial prejudice on the other.
METHOD
Participants
There were 230 participants, 138 women and 92 men, students at
Uppsala University and at the local authority-administered adult
education (in Swedish: Komvux), in the age range of 18 to 59 years
(M ˆ 23.1 years). They received cinema vouchers for their participation. The students represented various academic disciplines, such as,
social science, behavioral science, medicine, economics, technology,
and dentistry. The respondents participated also in the study by
Akrami et al. (2000).
Procedure and questionnaire
When they arrived, participants were given envelopes containing a
questionnaire of several parts. The part relevant to the present study
consisted of a preliminary set of items for the Modern and Classical
Sexism Scales, the Modern and Classical Racial Prejudice Scales
(Akrami et al., 2000), and a Conservatism Scale (constructed
following Ekehammar & Sidanius, 1982).
The Modern Racial Prejudice Scale ( ˆ 0.82) included items
such as ``Discrimination against immigrants is no longer a problem
in Sweden'' and ``Immigrants are getting too demanding in their
push for equal rights'', the Classical Racial Prejudice Scale
( ˆ 0.72) included items such as ``Immigrants do not keep their
homes tidy'' and ``Immigrants are generally not very intelligent'',
and the Conservatism Scale ( ˆ 0.74) included items such as
``Harder measures against criminals'' and ``Decreased social
welfare''.
The questionnaire items were embedded in items concerning
general subjects (e.g., ``Nuclear power is good'') in order not to
arouse participants' suspicion about the main purpose of the study.
Items were randomly mixed and half of the items were reversed.
Responses to all items were made on 5-point scales ranging from
strongly disagree (1) to strongly agree (5).
The experimenter gave oral instructions and told the participants
not to write names or make personal marks on the questionnaires in
order to ensure anonymity. When they finished the task, participants were told to put the envelope in a box placed in the room and
then received cinema vouchers. Participants were then thanked and
dismissed.
Classical and modern sexism
Scand J Psychol 41 (2000)
309
close to or lower than 0.05 indicate a satisfactory fit (Browne &
Cudeck, 1993). The Goodness-of-Fit Index (GFI) measures the
relative amount of variance and covariance jointly accounted for
by the model. The more variance accounted for by the model, the
better the fit. The GFI can range from 0 to 1, with higher values
indicating better fit (Bollen, 1989; JoÈreskog & SoÈrbom, 1993). The
additional three fit indices (NFI, NNFI, CFI) measure how much
better a model fits as compared to other models (Bollen, 1989).
These indices, too, can range from 0 to 1, with higher values
indicating better fit.
Scale development
Based on Swim et al.'s (1995) items, and Sears' (1988) classification system (denial of continued discrimination; antagonism
toward outgroups' demands; lack of support for policies designed
to help outgroups), 27 potential items were constructed to reflect
Swedish beliefs and ideas underlying classical and modern sexism.
We conducted a preliminary exploratory principal components
factor analysis of the items. The analysis yielded a two-factor
solution using a scree test for determining the number of factors.
All modern items loaded on the first factor, and most classical
items on the second. Items that loaded equally on both factors or
on the ``wrong'' factor were eliminated from subsequent analyses.
The 15 final items, 7 classical and 8 modern, are presented in
Appendix A.
RESULTS AND DISCUSSION
Exploratory factor analysis
LISREL analyses
The exploratory Principal Components factor analysis of
the final set of 15 Modern and Classical Sexism items
revealed a two-factor structure, where the Classical items
loaded on Factor 1 and the Modern items on Factor 2. The
internal consistency reliabilities were satisfactory ( ˆ 0.73
for the Classical and ˆ 0.80 for the Modern Sexism
Scale). The average inter-item correlation was r ˆ 0.31 and
r ˆ 0.36 for the Classical and Modern scales, respectively.
Factor loadings, item means, standard deviations, and
corrected item-total correlations are presented in Table 1.
As shown in Table 1, the modern items tend to have higher
item-total correlations and factor loadings than the
classical items.
Confirmatory factor analyses were carried out on the item
covariance matrices for the total sample using LISREL 8.14
(JoÈreskog & SoÈrbom, 1993). Three separate models were tested.
(1) The one-factor model tests if the covariance matrix of the modern
and classical items is best represented as one latent factor only.
(2) The uncorrelated two-factor model tests if the covariance matrix
can be represented as two (Modern and Classical) separate and
uncorrelated latent factors. (3) The correlated two-factor model tests
if the covariance matrix can be represented as two separate and
correlated factors.
The best factor solution was determined by using 2 tests.
However, because 2 values are influenced by sample size, five
additional indices were used to compare the goodness of fit of the
models. The Root Mean Square Error of Approximation
(RMSEA) measures the discrepancy per degree of freedom. Values
Table 1. Means, standard deviations, adjusted item-total correlations (r) and factor loadings for two-factor exploratory and correlated twofactor confirmatory analyses of the modern (M) and classical (C) sexism scales
Exploratory
Confirmatory
M
SD
r
M
C
M
C
Classical Sexism
1. Prefer a male boss.
2. Men's work is more important.
3. Women not talented.
4. No woman as prime minister.
5. Women bad at logical thinking.
6. Women care for children=old people.
7. Women bad drivers.
1.83
1.23
1.17
1.44
2.16
2.28
2.13
1.18
0.69
0.60
0.85
1.13
1.32
1.12
0.66
0.62
0.59
0.65
0.66
0.65
0.61
0.36
0.13
0.08
0.28
0.20
0.07
0.18
0.51
0.67
0.71
0.63
0.61
0.60
0.54
0
0
0
0
0
0
0
0.68
0.40
0.35
0.55
0.63
0.62
0.54
Modern Sexism
1. Discrimination of women no problem. a
2. Humiliation of women is unusual. a
3. Women treated in the same way. a
4. Too much of women's issues. b
5. Abolish Women's movement. b
6. Understand women groups. b
7. Don't adapt school to girls' needs. c
8. Equality in workplaces. c
1.95
2.13
2.09
2.20
2.16
2.50
3.34
1.87
1.03
1.22
1.21
1.24
1.19
1.14
1.21
1.02
0.77
0.52
0.50
0.78
0.76
0.66
0.52
0.73
0.75
0.44
0.45
0.74
0.74
0.69
0.43
0.76
0.28
0.09
0.07
0.31
0.25
0.11
0.11
0.18
0.82
0.44
0.44
0.99
0.86
0.69
0.43
0.74
0
0
0
0
0
0
0
0
Note. Scales range from 1 to 5, with lower scores indicating or recoded to indicate less racist responses. All rs are significant at p < 0.001.
Salient factor loadings are italicized. a Denial of continuing discrimination. b Antagonism toward demands. c Resentment about special
favors.
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B. Ekehammar, N. Akrami and T. Araya
Scand J Psychol 41 (2000)
Confirmatory factor analyses
The fit indices showed that the correlated two-factor model
fits the data relatively well (see Table 2, upper part).
Furthermore, the 2 tests indicated that the correlated
two-factor model gave a significantly better fit than the twofactor uncorrelated model, 2 (1, N ˆ 230) ˆ 82, p < 0.000.
The correlated two-factor model also yielded a better fit
than the one-factor model, 2 (1, N ˆ 230) ˆ 92, p < 0.000.
The correlation between the two factors was ˆ 0.68.
Further, the factor loadings for the correlated two-factor
solution were significantly greater than 0. The factor
loadings for the correlated two-factor model are presented
in Table 1 (right-hand part).
As hypothesized, the results support that classical and
modern sexism are distinguishable. The confirmatory factor
analyses indicated that the correlated two-factor solution of
modern and classical sexism items fitted the data significantly better than the one-factor and non-correlated twofactor solution.
Comparison of means
An analysis of the mean scores for the total sample showed
that the modern sexism scores (M ˆ 2.28, SD ˆ 0.75) were
significantly, t (229) ˆ 11.87, p < 0.001, higher than the
classical sexism scores (M ˆ 1.75, SD ˆ 0.63). Analysis of the
mean scores (ANOVA) for men and women showed that
men (M ˆ 2.60, SD ˆ 0.78) had significantly higher mean
sexism scores than women (M ˆ 2.07, SD ˆ 0.65) on the
Modern, F(1, 228) ˆ 29.87, ˆ 0.34, p < 0.001, as well as the
Classical Sexism Scale, F(1, 228) ˆ 22.99, ˆ 0.30, p < 0.001
(men: M ˆ 1.98, SD ˆ 0.74; women: M ˆ 1.60, SD ˆ 0.49).
These differences support the discriminant and construct
validity of the scales.
Correlations with other measures
In addition to examining gender differences, the construct
validity of the sexism scales was assessed by computing
product-moment correlation coefficients (r) with theoretically relevant measures. The analyses revealed that Modern
and Classical Sexism scores were positively correlated
( p < 0.000) with Modern Racial Prejudice (0.54 and 0.38,
respectively), Classical Racial Prejudice (0.30 and 0.39,
respectively), and Conservatism (0.34 and 0.28, respectively). The correlations were relatively similar to those
reported by other researchers working with conceptually
similar scales (e.g., Fazio et al., 1995; Ficarrotto, 1990; Glick
& Fiske, 1996; Pratto et al., 1994; Sidanius, 1993). With one
exception (the correlation between modern sexism and
conservatism was higher for men, 0.47, than for women,
0.20), there were no significant gender differences in the
magnitude of the coefficients. The highest correlations were
found between the modern versions of the sexism and racial
prejudice scales. The present correlations provide further
support for the construct validity of our sexism scales.
STUDY 2: CONFIRMING THE TWO-FACTOR
MODEL OF CLASSICAL AND MODERN SEXISM
The main purpose of Study 2 was to replicate the
confirmatory factor analyses conducted in Study 1, employing the same 17 items but another sample. Further, we
included a Swedish variant of the Social Dominance Scale
(Pratto et al., 1994) for a new construct validation of the
Modern and Classical Sexism Scales. Social dominance
orientation is an individual difference variable measuring
``one's degree of preference for inequality among social
groups'' (Pratto et al., 1994, p. 741). Measures of social
dominance orientation have been found to be highly
correlated with racial prejudice (Pratto et al., 1994), sexism
(e.g., Sidanius, 1993), and negative attitudes toward homosexuals (e.g., Whitley, 1999). Thus, we expect sexism to be
positively correlated with social dominance.
Table 2. Goodness of fit indices for three confirmatory factor models for the total group in study 1 and study 2
Model
2
df
2=df
GFI
RMSEA
NFI
NNFI
CFI
Study 1
One-factor
Uncorrelated two-factor
Correlated two-factor
233
223
141
90
90
89
2.59
2.48
1.58
0.863
0.891
0.922
0.094
0.076
0.053
0.774
0.784
0.864
0.820
0.832
0.934
0.846
0.857
0.944
Study 2
One-factor
Uncorrelated two-factor
Correlated two-factor
331
362
228
90
90
89
3.68
4.02
2.56
0.890
0.897
0.927
0.088
0.084
0.064
0.756
0.733
0.832
0.775
0.747
0.869
0.808
0.783
0.889
Note. GFI ˆ Goodness of Fit Index. RMESA ˆ Root Mean Square Error of Approximation. NFI ˆ Bentler's Normed Fit Index.
NNFI ˆ Bentler's Non Normed Fit Index. CFI ˆ Bentler's Comparative Fit Index. Values in bold indicate the best fit within the group.
# 2000 The Scandinavian Psychological Associations.
Classical and modern sexism
Scand J Psychol 41 (2000)
METHOD
311
Correlations with other measures
The procedure and the questionnaire were the same as in Study 1,
except that we added a Swedish translation of the Social Dominance
Scale (16 items). Sample items from this scale include ``Increased
social equality'' and ``Inferior groups should stay in their place''.
The reliabilities (Cronbach's alpha) of the different scales for the
present sample were satisfactory: Modern Sexism (0.74), Classical
Sexism (0.70), Modern Racism (0.83), Classical Racism (0.71), and
Social Dominance (0.87).
We computed, for validation purposes, product-moment
correlation coefficients (r) of scores on the Modern and
Classical Sexism Scales with scores on the Modern and
Classical Racial Prejudice Scales and the Social Dominance
Scale. The analyses revealed that Modern and Classical
Sexism scores were positively correlated ( p < 0.000) with
Modern Racial Prejudice (0.53 and 0.50, respectively),
Classical Racial Prejudice (0.34 and 0.52, respectively),
and Social Dominance (0.41 and 0.52, respectively). Except
for the correlation between the Modern Sexism Scale and
the Modern Racial Prejudice Scale, which was higher for
men (0.58) than for women (0.42), there were no significant
gender differences in the magnitude of the coefficients. The
correlations among the Sexism Scales and the Racial
Prejudice Scales were similar to those of Study 1. Further,
in line with the findings of previous research (Pratto et al.,
1994; Sidanius, Pratto & Bobo, 1996), our results confirm
that modern and classical racial and sexism scales share
common variance with the Social Dominance Scale.
RESULTS AND DISCUSSION
STUDY 3: STEREOTYPES OF WOMEN
Participants
There were 403 participants, 200 women and 203 men, students at
Uppsala University and at the local authority-administered adult
education (in Swedish: Komvux), in the age range of 18 to 52 years
(M ˆ 23.6 years). Respondents received cinema vouchers for their
participation. The students represented various academic disciplines, such as, social science, behavioral science, medicine,
economics, technology, and dentistry.
Procedure and questionnaire
Confirmatory factor analyses
As in Study 1, we tested three different models using
confirmatory factor analyses (LISREL). The fit indices
indicated that the correlated two-factor model fits the data
better than the one-factor solution (see Table 2, lower part).
Further, for the correlated two-factor solution, the factor
loadings were significantly greater than 0. The 2 tests
indicated that the correlated two-factor model fitted the
data significantly better than the uncorrelated two-factor
model, 2 (1, N ˆ 403) ˆ 134, p < 0.000. The correlated
two-factor model yielded also a better fit than the one-factor
model, 2 (1, N ˆ 403) ˆ 103, p < 0.000. Although highly
correlated ( ˆ 0.70), classical and modern sexism can be
distinguished. This is consistent with the results of Study 1
and those of Swim et al. (1995).
In Study 3, we investigate the content of people's knowledge
of cultural stereotypes of women. Following Devine's (1989,
Study 1) methodology, we examine whether (a) high- and
low-sexist people, and (b) men and women possess the same
knowledge of cultural stereotypes of women.
METHOD
Participants and procedure
The participants were the same as in Study 1. The part relevant to
this study contained one open-ended task. The instruction read
(translated from Swedish): ``In the space below, list at least five
common prejudices (in Swedish: foÈrdomar) of women in Sweden.
Note that we are not interested in your own personal beliefs, but in
those prejudices that are common in our society''.
RESULTS AND DISCUSSION
Comparison of means
Stereotypes about women
An analysis of the mean scores for the total sample showed
that the modern sexism scores (M ˆ 2.18, SD ˆ 0.70) were
significantly, t (402) ˆ 11.82, p < 0.001, higher than the
classical sexism scores (M ˆ 1.80, SD ˆ 0.62). Further, an
analysis (ANOVA) of the mean scores for men and women
revealed that men (M ˆ 2.40, SD ˆ 0.72) had significantly,
F(1, 401) ˆ 45.83, ˆ 0.32, p < 0.001, higher scores than
women (M ˆ 1.95, SD ˆ 0.61) on the Modern Sexism Scale
as well as on the Classical Sexism Scale, F(1, 401) ˆ 42.02,
ˆ 0.31, p < 0.001 (for men: M ˆ 1.99, SD ˆ 0.63, and for
women: M ˆ 1.61, SD ˆ 0.55).
Participants' responses were then content analyzed to form
different categories. If participants repeated one theme, it
was coded only once in that category. Coding of responses
was carried out by the first author and a fellow researcher.
This analysis resulted in 18 different categories (see Table 3).
To check the reliability of the coding, the second author
repeated the coding procedure independently of the first.
Both coders were blind to the participant's prejudice level.
According to the Kappa ( ) agreement coefficient (which
corrects for chance agreement), the intercoder reliability was
high ( ˆ 0.94).
# 2000 The Scandinavian Psychological Associations.
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B. Ekehammar, N. Akrami and T. Araya
Scand J Psychol 41 (2000)
Table 3. Proportion of total sample, men and women, and of high (HS) and low sexism (LS) participants mentioning each stereotype category
Rank
Category
Total
(n ˆ 230)
Men
(n ˆ 92)
Women
(n ˆ 138)
HS
(n ˆ 62)
LS
(n ˆ 62)
1
2
3
4
5
6
7
8
9
10
11
12
13
14
15
16
17
18
Miscellaneous negative
Better at taking care of children=old people
Emotional
Weak
Bad drivers
Not technical
Are made to do homework
Stupid
Miscellaneous neutral
Tell tales
Bad bosses
Bad at logical thinking
Dependent
Less educated
Miscellaneous positive
Bad at sports
Inferior to men
Have better simultaneous capacity
0.67
0.38
0.35
0.34
0.34
0.33
0.32
0.31
0.28
0.23
0.21
0.17
0.08
0.08
0.08
0.06
0.05
0.04
0.68
0.37
0.32
0.41
0.34
0.32
0.29
0.29
0.29
0.18
0.18
0.20
0.04
0.10
0.13
0.05
0.02
0.07
0.66
0.39
0.37
0.30
0.35
0.33
0.33
0.33
0.28
0.27
0.22
0.15
0.10
0.07
0.05
0.06
0.07
0.02
0.69
0.31
0.29
0.29
0.35
0.35
0.29
0.26
0.34
0.29
0.21
0.16
0.02
0.11
0.11
0.05
0.02
0.05
0.63
0.47
0.39
0.39
0.35
0.37
0.32
0.37
0.24
0.21
0.18
0.13
0.13
0.10
0.10
0.05
0.03
0.03
Significant differences between high- and low-sexist groups at p < 0.05.
Optimal extreme groups (cf. Crocker & Algina, 1986)
were created on the basis of participants' Modern Sexism
Scale scores: extreme low (lowest 27% of the sample,
N ˆ 62), mid-group (N ˆ 106), and extreme high (highest
27% of the sample, N ˆ 62). The means for the low- and
high-sexist groups were 1.45 (SD ˆ 0.19), and 3.26
(SD ˆ 0.45), respectively.
Of the 18 categories, only one (Dependent ) revealed a
significant difference between high- and low-sexist participants. Surprisingly, this category was mentioned more
frequently by low-sexist participants (see Table 3). However,
only one of the 18 categories (Miscellaneous positive) showed
a significant difference between men and women. This
category was mentioned most often by men. The heterogeneous category, Miscellaneous negative, was the most
mentioned category in the total sample as well as in the
various subgroups (men=women; high-in-sexism=low-insexism people). This category includes various negative
stereotypes about women, for example, vain, whining,
greedy, and disorganized.
GENERAL DISCUSSION
In Scandinavia, to our knowledge, no attempt has been
made by researchers to construct modern or classical sexism
scales. In the present paper, responding to the situation, we
addressed this issue by developing and testing a Modern and
a Classical Sexism Scale for a Swedish (Scandinavian)
context.
The confirmatory factor analyses in Study 1 and Study 2
support our hypothesis that modern and classical sexism are
distinguishable. The correlated two-factor model was
supported by the fit indices, which were higher than for
# 2000 The Scandinavian Psychological Associations.
Significant differences between men and women at p < 0.05.
the one-factor and uncorrelated two-factor models in both
studies. Interestingly, despite cultural differences, the results
in Study 1 and Study 2 are consistent with previous findings
in the USA (see Swim et al., 1995; Swim & Cohen, 1997).
However, because of the high correlation between the
modern and classical sexism factors, and the relatively small
differences in fit indices in some cases, one might argue that
modern and classical sexism are practically indistinguishable. Interestingly, a similar argument has been put forth
concerning modern vs. classical racism (see Dovidio,
Kawakami, Johnson, Johnson & Howard, 1997; Fazio et
al., 1995; Lambert, Cronen, Chasteen & Lickel, 1996;
Sniderman & Tetlock, 1986). However, following Akrami
et al. (2000), Sears (1988), and Swim et al. (1995), we argue
that the high correlations between the factors do not
necessarily lessen the importance of the distinction. Moreover, Mulaik, James, Van Alstine, Bennet, Lind, and Stilwell
(1989) argue that `` ... if two models applied to the same data
both obtain normed-fit indices in the 0.90s, the differences in
fit between them may indeed be small, involving only
differences in a few parameters, and yet the differences may
have considerable theoretical importance at a given historical moment'' (p. 434). Moreover, the analysis of the mean
scores for the total sample in Study 1 and Study 2 showed
that the modern sexism scores were significantly higher than
the classical sexism scores, revealing the blatant nature of
classical sexism. This can be considered as further support
for the theoretical importance of making this distinction.
Moreover, the gender differences in mean scores observed
in Study 1 and Study 2 support the construct validity of the
sexism scales. As expected, and consistent with previous
findings (Campbell et al., 1997; Glick & Fiske, 1996; Swim
et al., 1995), the results revealed that men scored higher than
Scand J Psychol 41 (2000)
women on both scales. The construct validity of the scales
was further tested by computing correlations with other
measures. In Study 1, as anticipated, the sexism scales were
positively correlated with modern and classical racial
prejudice as well as conservatism. In Study 2, in addition
to the correlations found between the racism and sexism
scales, we also obtained high correlations between the scales
of sexism and social dominance, a further confirmation of
the construct validity of our sexism scales. The relationships
obtained among racism, sexism, conservatism, and social
dominance are consistent with results reported by other
researchers (Ficarrotto, 1990; Glick & Fiske, 1996; Pratto et
al., 1994; Sidanius, 1993; Sidanius et al., 1996).
In Study 3, we examined people's knowledge of the
cultural stereotypes about women. The results revealed that
men and women, as well as high- and low-sexist people, did
not differ in their knowledge about cultural stereotypes.
Thus, akin to the findings of previous studies on racial
stereotypes (e.g., Devine, 1989), knowledge of the cultural
stereotypes concerning women seems to be much the same in
various subgroups.
Our studies add and extend the previous findings in the
sexism literature. From a practical point of view, our new
modern and classical sexism scales for a Scandinavian
context may save others from the work of developing these
measures. Because the Scandinavian countries share a
similar cultural background and have a common history
of equality between the sexes, it is reasonable to assume that
our scales could be profitably used in sexism research in the
other Scandinavian countries as well. More important, the
theoretical significance of our study lies in showing the
similarity of the factor structure with other studies
conducted in quite different cultural contexts. This is
interesting, because the Scandinavian countries have a
different history of equality between the sexes than, for
example, the USA (e.g., Glick & Fiske, 1996; Swim &
Cohen, 1997; Swim et al., 1995), Canada (Campbell et al.,
1997), and Germany (Eckes & Six-Materna, 1998). Finally,
to the best of our knowledge, no previous studies have
examined the content of the cultural stereotypes of women.
The present study has attempted to shed light on this issue.
Against this background, we suggest future research to
further investigate this theme.
This research was supported by Grant no. F0890=97 from the
Swedish Council for Research in the Humanities and Social
Sciences. We thank Andreas BirgegÊard for proofreading the
manuscript, and Dag SoÈrbom and Anders Holmberg for their
valuable suggestions concerning the LISREL analyses.
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APPENDIX A
Item Contents of the Classical and Modern Sexism Scales.
Classical Sexism
1. I prefer a male boss to a female.
2. A man's work is more important than a woman's.
3. Women are generally not very talented.
4. I would like to see a woman prime minister in Sweden.
5. Women are good at logical reasoning.
6. Women are better suited to look after children and old people.
7. Women are generally good drivers.
Modern Sexism
Denial of continuing discrimination
1. Discrimination of women is no longer a problem in Sweden.
2. Humiliating treatment of women in adverts is unusual.
3. Society treats men and women the same way.
Antagonism toward demands
4. The government puts too much emphasis on women's issues.
5. The women's movement serves no purpose and should be abolished.
6. It is easy to understand the viewpoints of women's groups.
Resentment about special favors
7. The school curriculum should be adapted to girls' needs.
8. Better measures should be taken to achieve equality (between the sexes) in workplaces.
Note. Items with an asterisk have reversed coding.
# 2000 The Scandinavian Psychological Associations.