International Journal of Sociology, vol. 43, no. 4, Winter 2013–14, pp. 63–86.
© 2013 M.E. Sharpe, Inc. All rights reserved. Permissions: www.copyright.com
ISSN 0020–7659 (print)/ISSN 1557–9336 (online)
DOI: 10.2753/IJS0020-7659430402
Aaron M. McCright and Aksel Sundström
Examining Gender Differences in
Environmental Concern in the Swedish
General Public, 1990–2011
Abstract: While numerous cross-sectional studies find modest gender differences in
environmental concern within the general publics of North American and European
countries, this pattern has not been examined over time—primarily due to a lack of
suitable data. Using twenty-two years of nationally representative survey data from
the Swedish general public, we test whether the theoretically expected relationship
between gender and environmental concern—where women are modestly more
proenvironmental than men—is robust over time. Results from our multivariate
ordered logistic regression models reveal a consistent pattern over the time period.
Across all available years of data, women report greater environmental concern
than men in the Swedish general public. Specifically, Swedish women report greater
worry about environmental destruction, greater worry about climate change, and
greater support for environmental protection than men. Thus, this gender difference
in environmental concern is indeed robust. The theoretically expected relationship
between gender and environmental concern is robust not only across environmental
concern indicators and countries of study but also over time.
A large body of scholarship in the past two decades finds that women in many
North American and European countries report moderately stronger environmental
concern (i.e., proenvironmental values, attitudes, beliefs, and behaviors) than do
men (Bord and O’Connor 1997; Dietz, Dan, and Shwom 2007; Flynn, Slovic, and
Mertz 1994; Hunter, Hatch, and Johnson 2004; McCright 2010; Stern and Dietz
1994; Stern, Dietz, and Kalof 1993; Xiao and McCright 2012b; Zelezny, Chua,
Aaron M. McCright is an associate professor in the Lyman Briggs College and Department
of Sociology at Michigan State University. Aksel Sundström is a doctoral candidate in the
Department of Political Science at the University of Gothenburg, Sweden.
63
64 international journal of sociology
and Aldrich 2000). Yet, with few exceptions (see, e.g., McCright 2010; Xiao and
McCright 2012a), nearly all of the studies documenting a gender difference in
environmental concern rely upon public opinion data from a single year. Dependence upon a single cross-section of a country’s general public is primarily due
to a lack of environmental concern indicators across a wide time series in existing
public opinion data sets. This limitation weakens our ability to better understand
the consistency or robustness of the gender difference in environmental concern
over time.
We aim to overcome this limitation by analyzing novel data from repeated
nationally representative cross-sectional samples of the Swedish general public
between 1990 and 2011. Given its lower level of gender inequality and less distinct
gender roles than are found in other countries (Rosenbluth, Salmond, and Thies
2006), the Swedish case represents a theoretically interesting context to investigate
gender differences in environmental concern. The yearly samples of the Swedish
population contain a few commonly used measures of environmental concern and
enjoy respectable response rates—from a low of 58 percent in 2008 to a high of 71
percent in 1992 (Vernersdotter 2012: 583). We employ multivariate ordered logistic
regression analyses on data over these twenty-two years to answer the following
research question: Is the theoretically expected relationship between gender and
environmental concern—where women are modestly more proenvironmental than
men—robust over time?
Gender and Environmental Concern
Research on environmental concern in many North American and European
countries in the past few decades consistently finds that women express slightly
greater environmental concern than men. This modest gender difference exists
whether environmental concern is operationalized via items measuring environment/economic trade-offs (e.g., McStay and Dunlap 1983), participation in proenvironmental activities (e.g., Hunter, Hatch, and Johnson 2004; Xiao and McCright
2012b), proenvironmental attitudes or an ecological worldview (e.g., Stern, Dietz,
and Kalof 1993; Xiao and McCright 2012a), or perceived seriousness of different
types of environmental problems (e.g., Mohai 1997; Xiao and McCright 2013).
The greatest gender differences are generally seen in studies dealing with the last
type of indicator—worry about specific environmental problems, especially those
local problems with clear health risks to family and community (e.g., Greenbaum
1995; Klineberg, McKeever, and Rothenbach1998; Mohai 1992).
Given that our main contribution here is to examine the relationship between
gender and environmental concern over time and because data limitations prevent
us from testing different hypotheses for why women are more proenvironmental
than men, we abstain from an extended literature review providing many details on
the major hypotheses explaining this gender difference in environmental concern.1
winter 2013–14 65
Instead, we briefly characterize this literature and identify those hypotheses enjoying
the most empirical support. We then discuss the relationship between gender and
environmental concern found in recent cross-national research, before describing
the Swedish context and reviewing the few relevant empirical studies of environmental concern in the Swedish general public.
Theoretical Explanations
In two classic pieces on gender and environmental concern in the general publics of Western societies, Davidson and Freudenburg (1996) and Blocker and
Eckberg (1997) each describe the prevailing explanations of gender differences
on environmentalism at the time. The two groups of explanations in these articles
have come to be known as “gender socialization” arguments and “gendered social
roles” arguments. Davidson and Freudenburg (1996) emphasize how childhood
socialization processes (e.g., Chodorow 1978; Gilligan 1982) lead males and
females to differ on important characteristics (e.g., concern about the safety and
care of others, value orientations, risk perceptions, trust in science) that correlate
with environmental concern.
Blocker and Eckberg (1997) focus on the influences of the social roles that
men and women differentially perform as adults (e.g., Greenbaum 1995). For the
most part, scholars focus on three productive or reproductive roles (employment
status, homemaker status, and parenthood) that presumably relate to environmental
concern. Much research finds that gender differences in environmental concern
are independent of the social roles and statuses that men and women differentially
occupy (e.g., McCright 2010; Mohai 1997). Over time, arguments on gendered
social roles have received little empirical support, especially when tested side-byside with gender socialization arguments.
Accordingly then, attention has turned to testing the explanations emphasizing
gender socialization. Among these, the safety concerns hypothesis (e.g., Blocker
and Eckberg 1997; Davidson and Freudenburg 1996; Xiao and McCright 2012a),
the values orientations hypothesis (e.g., Dietz, Kalof, and Stern 2002; Stern, Dietz,
and Kalof 1993), and the risk perceptions hypothesis (e.g., Bord and O’Connor
1997; Xiao and McCright 2012b) enjoy relatively consistent empirical support.
While this also was generally the case for the institutional trust hypothesis in earlier
decades (e.g., Blocker and Eckberg 1997; Davidson and Freudenburg 1996), recent
work suggests that this hypothesis no longer enjoys such empirical support (e.g.,
Xiao and McCright 2013).
Scholars typically employ the safety concerns hypothesis as a default explanation, invoking the claim that a feminine ethic of care is more strongly embodied
within women than men when a gender difference is found. A more rigorous test
of this hypothesis entails examining gender differences on attitudes about different
types of environmental problems: for example, local problems with clear health
66 international journal of sociology
and safety risks and global (or distant) problems with few or no perceived direct
health and safety risks to the respondents (see Xiao and McCright 2012a). In this
context, the safety concerns hypothesis expects a greater gender difference in attitudes about the first group of problems than about the second group.
The other two major hypotheses argue that a mediating factor—that is, risk
perceptions or values orientations—explain why women report greater environmental concern than men. Thus, a rigorous test of both hypotheses would involve
mediation analyses or structural equation modeling to properly model such a
mediating relationship. However, this approach also requires adequate measures
of such hypothesized mediating factors, which are often not found in existing data
sets. Beyond merely speculating, a less rigorous method to test the risk perceptions hypothesis is to examine the performance of gender in models predicting
environmental concern indicators that vary by how they tap into risk perceptions.
The risk perceptions hypothesis expects a greater gender difference on indicators
that tap risk perceptions than on those that do not. Those indicators that invoke
risk perceptions include measures of the perceived seriousness of a problem or
the amount a respondent worries about a problem. Those that do not seem to tap
into risk perceptions include measures of support for environmental protection and
general proenvironmental beliefs.
Gender Differences in Cross-National Research on Environmental
Concern
A growing number of cross-national analyses of environmental concern broaden
the focus on gender differences beyond North America and Western Europe to
different countries around the world. Scholars in this area either conduct separate
analyses for each country in their study (e.g., Hunter, Hatch, and Johnson 2004;
Kemmelmeier, Król, and Kim 2002; Lee and Norris 2000; Marquart-Pyatt 2007,
2008, 2012) or perform multilevel modeling on data pooled from many countries
(Franzen and Meyer 2010; Gelissen 2007; Givens and Jorgenson 2011; Hadler and
Haller 2011; Nawrotzki 2012).
Briefly, these studies find that women report stronger proenvironmental attitudes
(e.g., Franzen and Meyer 2010; Givens and Jorgenson 2011; Marquart-Pyatt 2007,
2008, 2012) and perform more private environmental behaviors (e.g., Hadler and
Haller 2011; Hunter, Hatch, and Johnson 2004) than men, though there is no gender difference in performance of public environmental behaviors (e.g., Hadler and
Haller 2011; Hunter, Hatch, and Johnson 2004). When environmental concern is
operationalized as willingness to pay or sacrifice to protect the environment, there
is either no gender difference (e.g., Gelissen 2007; Kemmelmeier, Król, and Kim
2002; Nawrotzki 2012) or men are more proenvironmental than women (e.g.,
Marquart-Pyatt 2008, 2012). Also, there does not seem to be a consistent gender
difference in environmental concern in Eastern Europe (Lee and Norris 2000; see
also Marquart-Pyatt 2007, 2008).
winter 2013–14 67
The Swedish Context
In a comparative perspective, the Swedish population does not seem to be either
very stable or remarkably extreme in its overall level of environmental concern.
In fact, environmental concern within the Swedish general public has fluctuated
significantly in past decades. For instance, incidents such as the massive deaths of
seals and the Chernobyl nuclear power plant disaster paved the way for increased
environmental concern in the late 1980s—ultimately resulting in emerging popular
support for the Green Party. Conversely, the economic crisis in Sweden in the early
1990s seems to have suppressed public support for environmental protection (see
Bennulf 1994, 1997; Harring, Jagers, and Martinsson 2009).
In the most thorough analysis of this matter over time, Harring, Jagers, and
Martinsson (2011) analyze why there have been “ups and downs” in environmental
concern within the Swedish general public in past decades. Studying nationally representative survey data from annual samples of the Swedish general public between
1987 and 2010, Harring, Jagers, and Martinsson (2011) find that the proportion of
the population that believes the environment is one of the most important problems
facing society has varied from its highest peak of 62 percent in 1988 to an all-time
low of 6 percent in 2003. In the years since, the proportion of the Swedish population that believes the environment is one of the most important problems facing
society has risen—now hovering around 20 percent.
With regard to gender and politics, the Swedish case is part of what is often
called “Scandinavian exceptionalism.” As with other Scandinavian countries,
Sweden has exceptionally high gender equality (e.g., Rosenbluth, Salmond, and
Thies 2006). For instance, for the past few decades, Sweden has been near the top
of rankings of countries around the world for several measures of gender quality,
such as the percentage of elected officials who are women (Inter-Parliamentary
Union 2006), and public perceptions of equality between men and women (Inglehart and Norris 2003).
The characteristics of the Swedish context we identified above make Sweden a
theoretically interesting case for investigating gender differences in environmental
concern. Given its lower levels of gender inequality and less distinct gender roles
than found in other Western countries, Sweden offers a “strong test” for finding
gender differences in environmental concern. Furthermore, the Swedish case is
not very different from other industrialized countries with respect to environmental concern in the general public. Given the observed fluctuation in aggregate
environmental concern in Sweden over the past few decades, finding a relatively
stable gender difference in environmental concern over the same period speaks to
the robustness of this effect.
Several studies find that women in the Swedish general public report stronger
environmental concern than men (Jagers 2006, 2009; Torbjörnsson, Molin, and
Karlberg 2011; Widegren 1998). Analyzing 1992–93 survey data from a nationally representative sample of the Swedish population, Widegren (1998) finds that
68 international journal of sociology
women are slightly more proenvironmental in their attitudes and behaviors than
men. Utilizing 2005 survey data from a nationally representative sample of the
Swedish population, Jagers (2006, 2009) reports that women are slightly more willing than men to pay higher taxes earmarked specifically to benefit the environment.
Investigating data from a 2009 survey of students in six different high schools in
Sweden, Torbjörnsson, Molin, and Karlberg (2011) find that female students report
greater biocentric values and are more inclined than male students to prefer preservation rather than utilization of nature. Analyzing nationally representative survey
data from 1986 to 1999, Bennulf (2000) illustrates that women are consistently
overrepresented among the supporters of the Green Party in Sweden.
Overall, the finding of a modest gender divide among citizens in the general
publics of North American and European countries is well established. Furthermore,
only a few, related theoretical explanations enjoy robust empirical support. Yet, since
the numerous studies documenting this modest gender divide on environmental
concern typically analyze public opinion data from only one year, we know much
less about how robust this pattern is over time—the topic to which we now turn.
The Study
To answer our research question about whether gender differences in environmental concern in Sweden are robust over time, we use nationally representative
survey data of the Swedish general public. The SOM Institute (Society, Opinion,
Media), an academic organization affiliated with the University of Gothenburg,
has administered a nationwide mail questionnaire annually since 1986. The SOM
surveys contain numerous questions on a wide selection of topics related to media
and politics. Because of data availability issues in the early years of this survey,
we focus solely on the repeated cross-sections between 1990 and 2011.2 The yearly
samples are representative of the Swedish population and enjoy respectable response
rates. The response rates for the 1990–2011 surveys range from a low of 58 percent
in 2008 to a high of 71 percent in 1992, and the sample sizes of the surveys during this time period range from a low of 1,573 in 1991 to a high of 5,007 in 2010
(Vernersdotter 2012: 583). The key survey items we use were administered to the
entire sample in some years and to randomly split subsamples in other years, so
the sample sizes we report in our tables of results are smaller.
Table 1 provides the name, coding, mean, and standard deviation for each of the
variables we use in our analyses. We operationalized our dependent variable, environmental concern, using three items found in many years across the time period of
this study. In each case, higher values mean greater environmental concern. Identical
or similar items are commonly used as measures of environmental concern (e.g.,
Bord and O’Connor 1997; Stern, Dietz, and Kalof 1993). As is common in large
nationally representative surveys, data availability limited us to utilize single-item
(rather than multi-item) indicators of environmental concern.
One environmental concern indicator, “support for environmental protection,”
1 (not at all worried) to 4 (very worried)
1 (not at all worried) to 4 (very worried)
1 (very bad proposal) to 5 (very good proposal)
0 (male) to 1 (female)
1 (low) to 3 (high)
1 (primary school) to 3 (college or university)
1 (18–29) to 4 (65 or older)
1 (never) to 4 (at least once a month)
1 (far left) to 5 (far right)
0 (does not support) to 1 (supports)
1 (no) to 3 (yes, very)
Coding
0.68
0.77
1.08
0.50
0.74
0.74
1.01
1.00
1.15
0.72
1.77
SD
3.38
3.25
3.38
0.50
2.03
2.06
2.48
1.75
3.03
Mean
*The original survey item measured the respondent’s annual household income with nine different categories. Due to increasing income levels
in Sweden, these categories changed between 1990 and 2011. Thus, the three different categories used here measure the relative income of respondents each year.
Worry about environmental destruction
Worry about climate change
Support for environmental protection
Gender
Income*
Education
Age
Religiosity (church attendance)
Political ideology
Party identification (9 dummy variables)
Devoted party supporter
Variable
Coding, Mean, and Standard Deviations for the Variables Used in the Study
Table 1
winter 2013–14 69
70 international journal of sociology
measures whether respondents believe it is “a very bad proposal” (1) to “a very good
proposal” (5) to “invest in an environmentally friendly society even if it means low
or no economic growth.” The other two indicators, “worry about environmental
destruction” and “worry about climate change” measure whether respondents are
“not at all worried” (1) to “very worried” (4) about environmental destruction and
climate change, respectively.
We measure gender dichotomously as the respondent’s sex: “male” (0) and
“female” (1). In our analyses we control for a group of social, demographic, and
political variables found to correlate with environmental concern (for an overview,
see McCright and Dunlap 2011). Briefly, we control for income (Van Liere and
Dunlap 1980), educational attainment (Greenbaum 1995; Heberlein 1981), age
(Pampel and Hunter 2012; Zelezny, Chua, and Aldrich 2000), religiosity (Clements,
McCright, and Xiao 2013; Sherkat and Ellison 2007), and political ideology and
party identification (Dunlap, Xiao, and McCright 2001; Fielding et al. 2012). To
accommodate both how income was initially recorded on the survey and increasing
income in Sweden over the time period, we measure income in relative categories:
“low” (1), “middle” (2), and “high” (3). For each year, these three categories represent the three lowest income categories, the three in the middle, and the three
highest income categories, respectively. Education ranges from “primary school”
(1) to “college or university” (3); age ranges from “eighteen to twenty-nine” (1)
to “sixty-five or older” (4).3
We measure religiosity as frequency of church attendance, ranging from “never”
(1) to “at least once a month” (4). We do not have 2004 data for this variable for
those respondents who were asked about their worry about environmental destruction and climate change, because religious attendance was not also asked on that
split sample survey (see Nilsson 2005). Political ideology is a scale from “far left”
(1) to “far right” (5). Sweden has a number of political parties, so we created nine
dummy variables to measure political party identification. Using identification with
the Center Party (Centerpartiet) as the reference category, we use eight dummy variables in our analyses to measure identification with the following parties (ordered
on the left–right continuum): Left Party (Vänsterpartiet), the Social Democrats
(Socialdemokraterna), the Green Party (Miljöpartiet de gröna), the Liberal People’s
Party (Folkpartiet liberalerna), the Christian Democrats (Kristdemokraterna), the
Moderates (Moderaterna), the Swedish Democrats (Sverigedemokraterna), and
other smaller political parties.4 Identification with the Swedish Democrats was
only included on surveys in 2007 onward. Finally, we measure how devoted each
respondent is as a party supporter, from “no” (1) to “yes, very” (3).
To answer our research question about whether the relationship between gender
and environmental concern is robust over the time period of our study, we ran
a series of multivariate ordered logistic regression models. For each dependent
variable, we ran a model with gender as a predictor, a model with gender and the
group of sociodemographic variables as predictors, and a model with gender, the
sociodemographic variables, and political variables as predictors. The complete
winter 2013–14 71
collection of models is available from the authors. Here, we present the results for
the full model for each dependent variable in each year for which data are available. To facilitate comparison across years and judge the relative importance of the
predictors, we present X-standardized odds ratios in our tables (Long and Freese
2005). An X-standardized odds ratio represents how much of a change in the odds
for each ordered comparison in the dependent variable is due to a standard deviation
increase in the predictor (Long 1997). As such, it allows for the direct comparison
of the effects of all predictors in the model.
Results and Discussion
Table 2 reports the percentages of men and women expressing environmental
concern across the twenty-two years of this study. There is indeed a statistically
significant gender difference in environmental concern across the entire time period.
In every year with only one exception, greater percentages of women than men
report environmental concern—regardless of its operationalization. Only for worry
about environmental destruction in 1991 is the gender difference not statistically
significant. In the pooled sample, 94.0 percent of women but only 86.0 percent of
men are pretty worried or very worried about environmental destruction, and 88.5
percent of women but only 77.5 percent of men are pretty worried or very worried
about climate change. Also, 53.1 percent of women but only 43.0 percent of men
believe that a proposal to invest in an environmentally friendly society (even if it
means low or no economic growth) is pretty good or very good. On average, there
is a slightly larger gender divide on worry about climate change (average yearly
percentage difference = 10.9 percent) and support for environmental protection
(10.4 percent) than on worry about environmental destruction (8.0 percent).
For the first environmental concern indicator (worry about environmental
destruction), the smallest percentage difference between men and women is 2.2
percent in 1991. Two more years from the 1990s also saw similarly small percentage differences (4.0 percent in 1994 and 2.9 percent in 1996). In the subsequent
years, the percentage differences were at least twice as large as these, with the
largest percentage difference of 11.7 percent in 2001. For the second environmental
concern indicator (worry about climate change), the smallest percentage difference
is 7.6 percent in 2007, and the largest percentage difference is 15.6 percent in
2010. For the third indicator (support for environmental protection), the smallest
percentage difference is 6.4 percent in 2007, and the largest percentage difference
is 14.3 percent in 1994.
We now turn to the results of our multivariate ordered logistic regression models
explaining worry about environmental destruction (Table 3), worry about climate
change (Table 4), and support for environmental protection (Table 5) over the time
period of this study. Most important, the pattern in Table 2, whereby women report
greater environmental concern than men in each year for each indicator, is confirmed.
Even when controlling for the group of social, demographic, and political predictors
72 international journal of sociology
Table 2
Percentages of Men and Women Espousing Environmental Concern in the
Swedish General Public, 1990–2011
Worry about
environmental
destruction
%“pretty” or “very”
worried about
environmental
destruction
Year
Men
Women
92.57
94.81
88.75
91.75
95.77*
95.77*
91.67
94.52*
Worry about climate
change
Support for
environmental
protection
%“pretty” or “very”
worried about climate
change
% believing a proposal
to invest in an
environmentally
friendly society is
“pretty” or “very” good
Men
Women
1990
1991
1992
1993
1994
1995
1996
1997
1998
1999
2000
2001
2002
2003
2004
2005
2006
2007
2008
2009
2010
2011
83.26
82.80
82.60
85.22
84.88
84.82
83.80
90.05
83.28
87.17
83.37
82.95
92.56*
94.54*
92.01*
94.20*
95.35*
94.71*
92.93*
96.91*
93.55*
94.11*
93.05*
92.73*
74.78
74.70
75.67
75.87
83.18
73.36
87.60
78.64
81.51
74.04
73.83
87.71*
83.36*
85.88*
84.20*
92.13*
86.00*
95.23*
91.98*
90.50*
89.61*
86.80*
Pooled
sample
85.97
94.03*
77.53
88.47*
Men
Women
42.35
46.65
52.18
53.20
44.63
41.37
53.77*
58.77*
60.66*
61.97*
58.91*
50.61*
39.97
39.87
41.89
36.69
40.75
40.28
39.92
40.00
37.00
36.73
53.19
45.43
48.65
50.15*
47.04*
50.50*
48.17*
52.88*
53.08*
52.27*
48.02*
49.40*
47.85*
59.60*
55.51*
58.21*
43.02
53.09*
*Percentage difference between men and women is statistically significant (α = 0.05).
winter 2013–14 73
described above, gender still has its theoretically expected effect on environmental
concern. We first discuss the results for gender in each of the three tables before
briefly discussing the performance of the other social, demographic, and political
predictors across the models.
Across our three tables, there are forty-six individual model years (sixteen years
in Table 3, eleven years in Table 4, and nineteen years in Table 5). While gender,
political ideology, and identification with the Green Party are the most powerful
predictors of environmental concern over the time period, gender is the only one
to have a statistically significant effect in all forty-six model years. Within the
Swedish general public over this time period, gender more consistently discriminates those with higher from those with lower environmental concern than any
other predictor.
Also notable are the magnitudes of the X-standardized odds ratios for gender
over the time period for each of the three environmental concern indicators. In most
years, gender is the second or third strongest predictor of environmental concern
(behind political ideology and Green Party identification). While the magnitudes
of these odds ratios do vary somewhat over the years, they do not do so monotonically. That is, in additional analyses not reported here, we find that the effect of
gender on each of the three environmental concern indicators has neither increased
nor decreased over the time period. Thus, to answer our guiding research question, the theoretically expected relationship between gender and environmental
concern—where women are modestly more proenvironmental than men—remains
quite stable over the time period.5
Figure 1 illustrates the net effect of gender on these three environmental concern indicators over the time period. The data series presented in this figure are
the standardized (log odds) effects of gender from the ordered logistic regression
models presented in Table 3 (Panel A), Table 4 (Panel B), and Table 5 (Panel C).
While the effect of gender does fluctuate slightly year to year (and this fluctuation is magnified because of the scale of the Y-axis), it is nevertheless relatively
stable over the time period—only varying across a range of a few tenths of a unit.
The effect of gender seems most stable in predicting support for environmental
protection (Panel C).
These results are consistent with the expectations of each of the three major
hypotheses predicting gender differences in environmental concern. Briefly, that
females in Sweden report greater environmental concern than their male counterparts is consistent with the explanation that women are socialized to have a stronger
ethic of care for others than are men (safety concerns hypothesis), that women have
a stronger altruistic value orientation than men (value orientations hypothesis), and
that women are more risk averse in their perceptions of environmental problems
than men (risk perceptions hypothesis). Yet, because of data limitations, we are not
able to adjudicate among these three hypotheses to determine which hypothesis
receives the strongest support.
Gender (female)
Income
Education
Age
Religiosity
Political
ideology
Left Party
Social Democrat
Party
Predictors
1.29*
0.97
1.08
0.78*
0.90
0.89
0.94
0.80
1.39*
1.00
1.19*
1.03
1.13*
0.80*
1.03
0.93*
1991–
2011 1991
0.67*
0.75*
0.87
1.29*
0.96
1.21*
0.96
0.95
1993
0.86
0.76*
0.98
1.29*
1.00
1.29*
0.92
1.07
1994
0.81
0.89
0.94
1.19*
0.93
1.13
0.81*
0.94
1996
0.78
0.76*
0.96
1.47*
1.03
1.26*
0.91
1.05
2000
0.98
0.77*
1.08
1.43*
0.93
1.29*
0.95
1.01
2001
0.66*
0.77*
0.92
1.29*
0.87*
0.95
1.12
1.13
2002
0.97
0.77*
1.00
1.37*
0.84*
1.19*
1.05
1.09
2003
1.08
0.84*
1.07
1.43*
0.98
1.08
1.09
2004
0.94
0.91
1.07
1.40*
0.99
1.14*
1.06
1.23*
2005
0.85
0.89
1.06
1.41*
1.01
1.19*
1.00
1.12
2006
0.72*
0.77*
0.95
1.28*
1.14
1.15*
1.00
1.26*
2007
0.83
0.75*
0.83
1.46*
1.09
1.08
1.01
1.23*
2008
0.87
0.78*
0.97
1.39*
1.04
1.21*
0.98
1.12*
2009
0.73*
0.77*
1.11
1.47*
1.03
1.26*
1.07
1.13*
2010
Results of Ordered Logistic Regression Models for Worry about Environmental Destruction in the Swedish General
Public, 1991–2011: X-Standardized Odds Ratios of Sociodemographic and Political Predictors
Table 3
0.83
0.84*
0.96
1.46*
1.07
1.20*
1.06
1.05
2011
74 international journal of sociology
0.89
0.96
0.72*
0.87*
0.99*
1.00
0.93*
0.92*
0.96
1.05*
0.99
0.87*
0.89
0.71*
0.81*
1.10
1.01
0.93
0.91
0.90
0.90
1.25*
1.00
0.97
0.90
0.77*
0.94
1.26*
1.11
0.91
0.93
0.75
0.87
1.14
1.05
1.19*
0.94
0.99
1.07
1.64*
0.99
0.92
0.82*
0.68*
0.85
1.10
1.04
0.96
0.96
0.92
0.96
1.40*
1.05
0.96
0.95
1.02
1.09
1.52*
0.98
0.96
0.84*
0.84
0.92
1.34*
1.09
0.88*
0.86*
0.78*
1.04
1.31*
1.04
0.87*
0.94
0.91*
0.79*
1.17*
0.93
0.79*
0.87
1.25*
0.79*
0.80
0.93
1.12
1.09
0.92*
0.86*
0.90
0.92
0.95
1.21*
1.03
0.87*
0.92
0.85*
0.81
0.91
1.30*
1.04
0.93*
1.06
0.99
0.85
0.96
1.27*
16,898 1,216 1,419 1,325 1,283 1,331 1,225 1,407 1,433 1,334 1,305 1,279 1,303 1,266 2,549 1,334 2,466
.05
.05
.03
.04
.04
.06
.06
.05
.05
.05
.05
.06
.04
.06
.05
.08
.06
2.04*
1.33*
Notes: The Swedish Center Party, “Centerpartiet,” is the reference category for the party dummy variables. No data on religiosity were available
for these respondents for 2004. Identification with the Swedish Democrat Party is only measured from 2007 onward. This variable is excluded
from the pooled sample. *p < 0.05.
N
Pseudo R 2
Green Party
Liberal People’s
Party
Christian
Democrat
Party
Moderate Party
Swedish
Democrat
Party
Other parties
Devoted party
supporter
winter 2013–14 75
Gender (female)
Income
Education
Age
Religiosity
Political ideology
Left Party
Social Democrat
Party
Green Party
Liberal People’s
Party
Predictors
1.46*
1.03
0.99
0.94
1.01
0.79*
0.92
0.86
1.13
0.84
0.99*
1.24*
0.99
2001
1.33*
0.96
1.07*
0.99
1.15*
0.85*
1.03
2001–
2011
0.94
0.87
1.13
1.23*
0.83*
0.92
1.04
1.19*
0.80*
1.02
2002
0.99
1.02
1.26*
1.26*
0.86*
1.07
0.99
1.07
0.85*
1.05
2003
0.97
0.93
1.28*
0.96*
1.12
1.31*
1.01
0.96
1.11*
2004
0.94
0.92
1.13
1.35*
1.00
1.06
1.02
1.30*
0.95
1.13
2005
1.09
1.08
1.33*
1.38*
0.96
1.11
0.92
1.08
0.99
1.20*
2006
0.95
0.81
1.17
1.18*
1.08
1.10
0.93
1.34*
0.81*
1.10
2007
0.82*
0.83
1.33
1.41*
1.03
1.05
1.00
1.33*
0.83*
0.78*
2008
0.88*
0.82
1.13*
1.33*
1.01
1.09
0.97
1.10*
0.87
0.91
2009
0.91
0.78
1.23*
1.44*
1.02
1.11
.0.99
1.15*
0.77*
0.98
2010
Results of Ordered Logistic Regression Models for Worry About Climate Change in the Swedish General Public,
2001–2011: X-Standardized Odds Ratios of Sociodemographic and Political Predictors
Table 4
0.98
0.87
1.17*
1.43*
0.99
1.08
0.99
1.09*
0.84*
0.93
2011
76 international journal of sociology
0.99
0.88
0.96
1.02
1,407
.03
0.96
0.94
0.89
1.03
1,225
.03
1.05
1,433
.02
0.96
0.99
0.93
1.14*
1,334
.02
0.94
0.94
0.92
1.00
1,305
.04
0.96
0.91
0.80*
1.17*
1,279
.04
0.93
0.99
0.89
1.05
1,266
.06
0.84*
0.94
0.92
0.90
1.08
1,303
.04
0.80*
0.72*
0.87
0.90
1.11*
2,549
.04
0.92*
0.85*
0.83*
0.81*
1.00
1,334
.06
0.87*
0.91
0.83*
0.78*
1.14*
2,466
.04
0.94*
1.11
0.95
0.87
Notes: The Swedish Center Party, “Centerpartiet,” is the reference category for the party dummy variables. No data on religiosity were available
for these respondents for 2004. Identification with the Swedish Democrat Party is only measured from 2007 onward. This variable is excluded
from the pooled sample. *p < 0.05.
Christian
Democrat
Party
0.95*
Moderate Party
0.95
Swedish
Democrat
Party
Other parties
0.96*
Devoted party
supporter
1.06*
N
15,567
Pseudo R 2
.03
winter 2013–14 77
1.06
0.97
1.05
1.07
1.10* 1.08
1.12*
1.04
1.03
0.84
0.97
0.77* 0.92
0.86
0.97
0.95
1.01
0.87* 0.88
0.77* 0.81* 0.76
0.95
0.99
0.94
1.03
0.93
0.90
1.05
0.95
0.87
1.11
0.95
0.95
0.93
0.80* 0.83
1.06
0.94
0.78*
1.08
0.90* 0.75* 0.76* 0.79* 0.79* 0.95
0.73* 0.85
0.94
Liberal
People’s
Party
0.63* 0.72* 0.62* 0.83
0.86* 0.95
0.71* 0.72* 0.76* 0.81* 0.74* 0.69* 0.76* 0.76* 0.73* 0.74* 0.75* 0.80* 0.71*
0.96
0.81* 0.62* 0.62* 0.83
0.88
1.07
1.07
1.12* 1.11* 1.15* 1.13* 1.09
1.14* 0.98
1.03
1.36* 1.22* 1.62* 1.64* 1.23* 1.40* 1.17* 1.20* 1.37* 1.40* 1.45* 1.57* 1.47* 1.38* 1.48* 1.37* 1.44* 1.40* 1.43* 1.27*
0.98
1.04
1.05
0.89* 0.88* 0.90* 1.03
1.01
1.20* 1.17* 1.09
1.10
0.92
Green Party
0.96
1.08
1.03
1.13* 1.10* 1.20* 1.04
0.97
1.02
Social
Democrat
Party
0.97
0.74* 0.70* 0.76* 0.81* 0.73* 0.68* 0.74
0.95* 0.83* 0.86
1.13* 1.13* 1.00
Left Party
1.14* 1.14* 1.09
1.00
0.98
0.92* 0.93
1.13* 1.11
Political
ideology
0.97
1.18* 0.99
1.02
0.90* 0.90* 0.93
Religiosity
0.96
0.93
1.06* 1.18* 1.16* 1.18* 1.17* 1.17* 1.11
0.98
0.91
Age
1.18* 1.04
0.93
Education
0.88* 0.91
1.04* 0.96
Income
1.10
1.20* 1.25* 1.31* 1.17* 1.20* 1.27* 1.24* 1.23* 1.18* 1.16* 1.20* 1.24* 1.19* 1.21* 1.18* 1.26* 1.22* 1.15* 1.17* 1.19*
0.92* 0.88* 0.91
Gender
(female)
1990–
2009 1990 1991 1992 1993 1994 1995 1997 1998 1999 2000 2001 2002 2003 2004 2005 2006 2007 2008 2009
Predictors
Results of Ordered Logistic Regression Models for Support for Environmental Protection in the Swedish General Public,
1990–2009: X-Standardized Odds Ratios of Sociodemographic and Political Predictors
Table 5
78 international journal of sociology
1.07
.04
.06
.06
.04
.03
.05
0.89* 0.95
0.88* 0.98
1.03
0.93
1.04
1.00
.05
.05
.04
1.04
0.99
1.02
1.00
0.97
1.00
1.02
1.11
0.95
1.01
0.96
0.97
0.95
1.01
1.08* 1.01
0.90
0.92* 0.95
0.92* 0.95
0.95
1.14*
0.91*
0.89*
0.79* 0.87
0.90* 0.98
.04
.04
.03
.05
.04
.04
.04
.05
.04
.04
.05
2,504 2,698 2,552 1,404 1,429 1,394 2,686 2,636 2,734 1,341 2,679
1.05
0.99
0.94
0.77* 0.87* 0.86* 0.89
0.96
Notes: The Swedish Center Party, “Centerpartiet,” is the reference category for the party dummy variables. No data on religiosity were available
for these respondents for 2004. Identification with the Swedish Democrat Party is only measured from 2007 onward. This variable is excluded
from the pooled sample. *p < 0.05.
Pseudo R 2
0.98* 0.93
1.05
1.02
0.89* 0.99
36,3191,148 1,216 1,536 1,419 1,325 1,424 1,292 2,02
1.03* 0.99
Devoted
party
supporter
N
0.97
0.64* 0.72* 0.72* 0.78* 0.69* 0.82* 0.72* 0.83
0.94
0.87
1.02
Other parties 0.95* 0.74* 0.88* 0.89* 0.83* 0.89
0.99
0.79* 0.61* 0.60* 0.72* 0.68* 0.84
Swedish
Democrat
Party
0.95
Moderate
Party
0.89
0.93* 0.96
Christian
Democrat
Party
winter 2013–14 79
1
B 1 1988
1988
1993
1993
1998
1998
2003
2003
2008
2008
Worry about climate change
2013
2013
Notes: The data series presented in this figure are the standardized (log odds) effects of gender from the ordered logistic regression models
presented in Table 3 (Panel A), Table 4 (Panel B), and Table 5 (Panel C).
C
Support for environmental destruction
1 11988
1993
1998
2003
2008
2013
1988 1993 1998 2003 2008 2013
1.11.1
1.21.2
1.31.3
1.41.4
1.51.5
1.61.6
1
A 11988
1988
1993
1998
2003
2008
2013
1993
1998
2003
2008
2013
1.1
1.1
1.11.1
1.3
1.3
1.4
1.4
1.5
1.5
1.6
1.6
1.2
1.2
Worry about environmental destruction
1.21.2
1.31.3
1.41.4
1.51.5
1.61.6
Figure 1. Net Effects of Gender on Environmental Concern in the Swedish General Public, 1990–2011
80 international journal of sociology
winter 2013–14 81
The effects of the social, demographic, and political variables in our full
multivariate ordered logistic regression models are also notable. Besides gender,
political ideology is also a relatively robust predictor. Consistent with much existing
research (e.g., Dunlap, Xiao, and McCright 2001), Left-leaning individuals report
greater environmental concern than their right-leaning counterparts in thirty-eight
model years. Also expected given past research (e.g., Jones and Dunlap 1992),
educational attainment has a positive effect on environmental concern, but only in
fourteen model years (and none for worry about climate change in Table 4). The
positive effect of religiosity on environmental concern in twenty-five model years
is somewhat unexpected given that most studies find no relationship between religiosity and environmental concern (e.g., Eckberg and Blocker 1996; Hayes and
Marangudakis 2000).
Two other sociodemographic variables (age and income) are not robust predictors of environmental concern. Respondents making higher income report lower
environmental concern than their respective lower income counterparts in twelve
model years. Younger adults report greater worry about environmental destruction
than their older counterparts in two model years, but older adults report greater worry
about climate change and greater support for environmental protection than their
younger counterparts in eight model years. We do acknowledge that the effect of
our age variable is ambiguous in that it may mask period or cohort effects—which
can be addressed in future work.
Finally the performance of the dummy variables measuring party identification
(with the Center Party as the reference category) deserves attention. The magnitude
and direction of these effects are broadly in accordance with expectations. Identification with the Green Party (compared to the Center Party) is associated with
greater environmental concern in thirty-seven model years. Identification with rightleaning parties is associated with lesser environmental concern, but in fewer model
years: twenty-two for the Moderates and twelve for the Christian Democrats. Past
research demonstrates that the Swedish general public perceives the right-leaning
Center Party as significantly more proenvironment than the two leftist parties (the
Left Party and Social Democrats), because of the former’s history of opposition
to nuclear power (Kitschelt 1986; Löfstedt 1996; Oscarsson 1998). Thus, it is not
surprising that identification with these two leftist parties (compared to the Center
Party) is not associated with greater environmental concern.
Conclusion
Numerous studies of the general publics in North American and European countries
find that women report modestly stronger environmental concern than men. Yet,
little research has examined the relationship between gender and environmental
concern over time to determine how temporally robust this relationship is. We
aimed to increase our knowledge of whether the relationship between gender and
environmental concern is robust over time by analyzing a public opinion data set
82 international journal of sociology
containing twenty-two years of survey data from the Swedish general public.
We argued that Sweden is a theoretically interesting case. Given its lower levels of
gender inequality and less distinct gender roles than are found in other Western countries, Sweden is a “strong test” for finding gender differences in environmental concern.
Furthermore, given the noticeable fluctuation in aggregate environmental concern in
Sweden over the past few decades, finding a relatively stable gender difference in environmental concern over the same period speaks to the robustness of this effect.
Controlling for the effects of a group of sociodemographic and political variables,
our multivariate ordered logistic regression models demonstrate that women report
greater environmental concern than do men in each year of our study. These results
are consistent with the expectations of the safety concerns, risk perceptions, and
values orientations hypotheses, though data limitations prevent us from directly
testing each explanation. The net effect of gender is relatively stable over the years,
despite some expected year-to-year fluctuation. Thus, while other studies find that
the theoretically expected relationship between gender and environmental concern
is robust across countries and environmental concern indicators, we further demonstrate that it is robust over time.
Future research on this topic should examine whether the relationship between
gender and environmental concern is consistent over time not only in other European and North American countries but also in countries in Asia, Africa, and
South America. For this to occur, we need improvements in our social science data
infrastructure that would widely benefit the entire scholarly community. Annual
national surveys on the environment, which include sufficient items to measure
gender, environmental concern, and other key variables with composite indicators,
would serve as a valuable community resource. Such surveys could utilize repeated
cross-section samples as well as panel subsamples to examine change not only over
time but also within individuals. This community resource, which would allow us to
answer a wide range of research questions in environmental social science, would
be especially beneficial for scholars studying the relationship between gender and
environmental concern.
Notes
1. Interested scholars can read the notable literature reviews in this area (e.g., Blocker
and Eckberg 1997; Davidson and Freudenburg 1996; McCright 2010).
2. An alternative assessment of the robustness of the gender effect over time involves
examining the effect of gender over the life course of individuals; however, panel data for
such analyses are not available.
3. Pampel and Hunter (2012) recently demonstrated how to disentangle age, period,
and cohort effects when examining environmental concern in the United States. Because
of the way age was originally measured in the SOM surveys and our emphasis on gender,
we have chosen not to investigate age, period, and cohort effects.
4. Voters as well as experts generally place the major political parties in Sweden on
the left–right continuum in the following order: the Left Party, the Social Democrats, the
Green Party, the Center Party, the Liberal People’s Party, the Christian Democrats, the
winter 2013–14 83
Moderates, and the Swedish Democrats (see, e.g., European Election Database 2010;
Statistics Sweden 2010; Volkens et al. 2012). Scholars also discuss how Swedish political
parties are associated with environment-related policy issues. This research suggests that
the parties can be placed on a “green” continuum, which does not align perfectly with
the left–right continuum (e.g., Gilljam and Holmberg 1995). Oscarsson (1998) analyzes
this at length, studying the perceptions of voters with nationally representative data from
the mid-1990s. Voters in Sweden perceive the Green Party as the most proenvironmental
party. While this comes as no surprise, they also perceive the right-leaning Center Party
as significantly more proenvironmental than the two leftist parties: the Left Party and
the Social Democrats (Oscarsson 1998: 21). This partly stems from the Center Party’s
mobilizing to oppose nuclear power in the 1970s and 1980s (Kitschelt 1986; Löfstedt
1996). The Liberal People’s Party and the Moderates are considered by voters as the least
proenvironmental parties (Oscarsson 1998: 21). The National Election Study of 2010 finds
a similar pattern in an analysis of which political issue is the most important for Swedish
voters when deciding which political party to vote for in the national legislative elections.
The highest proportion of people who state that the environment is the most important
political issue is found among citizens voting for the Green Party (57 percent) followed by
those voting for the Center Party (20 percent). Correspondingly, the lowest proportion is
found among voters of the Moderates (5 percent) (Statistics Sweden 2010: 61).
5. While we acknowledge Flood’s (2009: 68) point that comparing odds ratios across
ordered logistic regression models should be done with care since “unobserved heterogeneity
can vary across the compared samples, groups, or points in time,” we are still confident that
our obtained results show the relative robustness of gender over time.
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