This supports our assumption of no selection into having at least one daughter, conditional on family size. These are dummy variables taking a value of one if there are only daughters of the respective age group in the household (columns  and ), if the oldest child in the household is female and in the respective age group (columns  and ), and if there is at least one daughter with the age group based on the oldest daughter (columns  and ), respectively, and zero otherwise. For example, feminist economist Stephanie Seguino uses a definition which is very close to that of Pearse and Connell, describing them as ‘acceptable behavioural boundaries for men and women, congruent with the gender division of labour and male power’ (2007, p.2).
We then turn to examine attitudes among women. While this variable relates less directly to attitudes towards the gender division of work, it still captures attitudes towards gender norms more generally.
Attitudes towards gender norms seem to be malleable to experiences during adulthood such as parenting a daughter, thus suggesting that indirect exposure to disadvantage has the potential to change people’s attitudes. August 30, 2009: Brooks Raley assigned to Boise Hawks from AZL Cubs.
Our main findings are reflected with this alternative outcome variable: parenting daughters makes men less traditional, the effect is strongest for fathers of school-age daughters, and there are no statistically significant effects for mothers that are robust to the introduction of individual FEs.28 Finally, we find that the main results with the binarized dependent variable (see columns  and  of Tables 1 and 2) are robust to estimating logit and FE logit models (Table A.9).29. Our main regressor of interest is the binary variable ‘at least one daughter’ that takes on a value of one if the respondent has at least one daughter living in the household, and zero otherwise.14 To account for the potential interactive effect between child gender and child age on attitudes to gender norms, we also distinguish different age groups: ‘daughter 0 to 5’, ‘daughter 6 to 10’, and ‘daughter 11+’ are dummy variables indicating whether there is at least one daughter of the respective age group living in the household. We recoded the answers so that higher values mean stronger agreement with the statement, hence more traditional attitudes.12 We also binarize the dependent variable into non-traditional (taking a value of zero for ‘disagree’ or ‘strongly disagree’) and traditional (taking a value of one for ‘neither agree nor disagree’, ‘agree’, and ‘strongly agree’) and report results alongside those for the ordinal dependent variable.13. }); In addition to performing these robustness checks, we note that FEs account for the bias arising from time-invariant unobservable characteristics that are correlated with both the probability to live with daughters and with attitudes. It shows the mean value of the ordinal (panel a) and binarized (panel b) outcome variable.
Such a test is described in Fitzgerald et al.
} We find that there is a negative but not statistically significant association between having a first daughter and the total number of children for both men and women.25 However, we do find a negative and statistically significant correlation between having at least one daughter among the two oldest children and the number of total children, for the subsample of individuals who have at least two children by the last wave.26 This implies that those who have two sons are more likely to have further children compared to those with at least one daughter and hence there is positive selection of having at least one daughter.
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