Green returns to education: Does education affect pro-environmental attitudes and behaviors in China?

individuals that are more affected by the CSL have higher educational attainment. The F-test statistics of the first-stage regressions are 129.41 and 143.44 respectively, which exceed the weak instrument critical value established by Staiger and Stock [43]. Taken together, these results indicate that the weak identification is unlikely an issue in our study. More importantly, education remains highly significant in the second-stage regression. The magnitude of the coefficient is not negligible. Examining the economic significance of the education coefficient in column (1), we find that an extra year of schooling is associated with an increase of 0.012 in the pro-environmental attitudes, which is equivalent to 2.1% (= 0.012/0.569) of the sample mean. The coefficient estimate in column (2) implies that an additional year of schooling results in an increase of 0.011 in the pro-environmental behaviors, which is approximately 2% (= 0.011/0.539) of the corresponding sample means. Therefore, the impact of education on pro-environmental attitudes and behaviors is economically sizable on annual basis. Our results show that the IV estimates are larger than the corresponding OLS estimates for regressions of pro-environmental attitudes and behaviors. The downward bias of OLS estimates might be driven by measurement error in survey response. For example, some studies suggest that mea- surement errors in schooling can lead to downward bias in the OLS estimate of the effect of schooling [29,44]. 4.2. Heterogeneity analysis

The benchmark results show that educational attainment has strong and positive impacts on the pro-environmental attitudes and behaviors of the whole sample. However, these impacts might differ across different groups of people. In this section, we examine the heterogeneity in the effects of education in the following three aspects. Specifically, we split our sample based on a given individual characteristic (e.g., gender) and then within each subsample, regress pro- environmental attitudes and behaviors on education, controlling for other observable characteristics. First, we examine whether the education effects on pro-environmental attitudes and behav- iors differ by gender. In Table 5, the results in columns (1) and (7) are for male group, while those in columns (2) and (8) are for female group. For pro-environmental attitudes, the esti- mated coefficient on education is significant for males, but insignificant for females, suggesting that men are more likely to have pro-environmental attitudes through education. This might Table 4. The effects of education on pro-environmental attitudes and behaviors, 2SLS results. Pro-environmental attitudes Pro-environmental behaviors (1) (2) Education 0.012�� 0.011�� (0.005) (0.004) Controls Yes Yes Observations 3,456 3,661 First-stage estimation CSL 2.937��� 3.141��� (0.258) (0.262) F-statistics 129.41 143.44 Note: The table reports the estimates from 2SLS regressions of education on pro-environmental attitudes and behaviors. Columns (1) and (2) report the results for pro-

environmental attitudes and for pro-environmental behaviors, respectively. All regressions have the same controls as in column (2) of Table 3, but their coefficients are suppressed for brevity. Robust standard errors are clustered at the cohort-city level and are reported in parentheses. ���, �� and � indicate 1%, 5% and 10% significance levels, respectively. https://doi.org/10.1371/journal.pone.0263383.t004 PLOS ONE Green returns to education PLOS ONE | https://doi.org/10.1371/journal.pone.0263383 February 3, 2022 12 / 20