This study tests theoretical arguments about gender differences in scientific knowledge and environmental concern using 8 years of Gallup data on climate change knowledge and concern in the US general public. Contrary to expectations from scientific literacy research, women convey greater assessed scientific knowledge of climate change than do men. Consistent with much existing sociology of science research, women underestimate their climate change knowledge more than do men. Also, women express slightly greater concern about climate change than do men, and this gender divide is not accounted for by differences in key values and beliefs or in the social roles that men and women differentially perform in society. Modest yet enduring gender differences on climate change knowledge and concern within the US general public suggest several avenues for future research, which are explored in the conclusion.
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I use climate change and global warming interchangeably, although the former technically connotes all forms of climatic variability introduced by the warming of Earth’s surface and oceans stemming from the increased accumulation of greenhouse gases in Earth’s atmosphere. The increased concentration of such gases strengthens the natural “greenhouse effect” whereby the atmosphere absorbs the sun’s radiation rather than allowing it to escape into space (see National Research Council 2001, 2008).
Since this study does not address environmental behavior, I do not review the large literature on the relationship between gender and environmental behavior (see, e.g., Hunter et al. 2004). This review also is limited to the association between gender and environmental concern in the United States.
This is consistent with the finding that women are more concerned than are men about a wide range of risks—not just environmental ones (e.g., Slovic 2001).
Indeed, at least one study finds that general environmental values and beliefs are the strongest correlates of climate change attitudes and beliefs (Kellstedt et al. 2008).
The other two are as follows. According to the “Trust in Science and Technology Hypothesis” (Blocker and Eckberg 1997) or the “Institutional Trust Hypothesis” (Davidson and Freudenburg 1996), some argue that differences between men and women in levels of trust in science and technology explain gender differences in environmental concern. Many studies find that women tend to trust science and technology less than do men (e.g., Blocker and Eckberg 1997; Davidson and Freudenburg 1996; Flynn et al. 1992; Fox and Firebaugh 1992), and others find that trust in science and technology is negatively related to environmental concern (e.g., Freudenburg 1993; see review in Davidson and Freudenburg 1996). Also, Stern et al. (1993) argue that differences in men’s and women’s value orientations explain gender differences in environmental concern. Indeed, the authors provide evidence that women’s significantly strong embrace of altruism is the basis for gender differences in environmentalism (Dietz et al. 2002).
Gallup interviewers begin each telephone interview with well-established questions on a range of general topics before turning at the end of their interviews to specific questions on environmental issues. The 2001 poll has a sample of 1060 adults interviewed between March 5–7; the 2002 poll has a sample of 1006 adults interviewed between March 4–7; the 2003 poll has a sample of 1003 adults interviewed between March 3–5; the 2004 poll has a sample of 1005 adults interviewed between March 8–11; the 2005 poll has a sample of 1004 adults interviewed between March 7–10; the 2006 poll has a sample of 1000 adults interviewed between March 13–16; the 2007 poll has a sample of 1009 adults interviewed between March 11–14; and the 2008 poll has a sample of 1012 adults interviewed between March 6–9.
As is typical in most national surveys, the Gallup Organization employs weighting procedures on the sample data to ensure that the samples are representative of the American adult population. I do not employ data weights when performing multivariate analyses, because weighting can lead to inflated standard errors and misleading tests of significance (Winship and Radbill 1994).
The lack of a statistically significant temporal trend in the key dependent variables (i.e., the climate change knowledge and concern indexes) indicates that pooling is appropriate.
For both variables, I calculated an unweighted mean for the pooled sample before creating a centered score (raw score minus mean).
Global warming ranks relatively low on lists of environmental problems citizens worry about. For instance, in 2008, global warming ranked tenth out of twelve environmental problems (above urban sprawl and acid rain) (Jones 2008). For the most part, Americans worry much more about local air and water pollution problems than they do about global problems (such as the loss of tropical rain forests, damage to the earth’s ozone layer, and global warming).
The results of this model predicting scores on the climate change knowledge index are similar to the results of separate models predicting values of each of the three individual climate change knowledge items. For space reasons, I only present the results of the former model.
To check for possible multicollinearity problems, I examined the variance inflation factors (VIF) from each of the models in Table 5. The greatest VIF in Table 5 is 1.49 in the fully specified model, well below the threshold of 10 that is cause for concern about multicollinearity (see Chatterjee et al. 2000).
The sign and magnitude change for the religiosity coefficient in the fully specified model bears some discussion. The statistically significant, positive coefficient of religiosity in the fully specified model is opposite of the zero-order correlation between religiosity and the climate change concern index (Pearson’s r = −.097; N = 4078; p < .001). I examined a series of partial correlations between religiosity and climate change concern to explore this shift. Controlling for both climate change knowledge and political ideology switches the negative correlation between religiosity and concern to positive (Pearson’s r = .053; N = 3072; p = .003). In other words, the effect of religiosity net of the effects of climate change knowledge and political ideology is positive (opposite of what H5a expects), which is further evidence to reject this hypothesis.
The positive coefficient on homemaker status does achieve statistical significance in the fully specified model. Nevertheless, running even a few variations of the fully specified model with one or more of the key variables removed suggests that this statistically significant, positive coefficient on homemaker status is not sufficiently robust to offer support for H7.
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Thanks are extended to Riley E. Dunlap and the Gallup Organization for making the data available for analysis. The author also thanks Chenyang Xiao for his helpful advice. The author is grateful to the reviewers for their productive feedback.
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McCright, A.M. The effects of gender on climate change knowledge and concern in the American public. Popul Environ 32, 66–87 (2010). https://doi.org/10.1007/s11111-010-0113-1
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