and FB families, we first show that our results survive controlling for indicators for the number of children (column (2) of Table 5) and controlling for the proportion of sons (Table 6).<sup>34</sup> Second, we explicitly estimate the effect of expected dowry on differential fertility and sex-selection behavior in FG relative to FB families. To do so, we use data on fertility and sex-selection from the 199899 and 2005-06 rounds of the Demographic Health Survey of India, also known as the National Family Health Survey (NFHS).<sup>35</sup> We restrict our analysis to the rural sub-sample of NFHS since our dowry data from REDS does not cover urban India. We prefer using NFHS for this analysis as it reports complete birth histories, including the dates of live births and of any child deaths, for each surveyed woman, where as in REDS we need to impute birth year and birth order from children’s ages. Moreover, NFHS is substantially larger in sample size than REDS.

We estimate the following equation by using the number of children and the proportion of sons among second and higher parity births at the time of survey for mother _i_ from caste _c_ in state _s_ and whose first child was born in year _t_ as the dependent variables: The vector **X** _i_ comprises parents’ years of schooling, parents’ ages, and indicators for religion, month of survey, and household standard of living at the time of survey. The inclusion of the FG main effect allows us to control for any changes in fertility or sex-selection due to firstborn sex that could result from factors unrelated to dowry; for instance, higher fertility among FG families due to the desire for at least one son. If certain castes are richer and therefore can more easily afford sex-selection technology, the caste fixed effects ( _ωc_ ) would capture this confounding variation. If there is state or time variation in wealth status of various castes that can explain the differential fertility behavior, it would be captured by the state-caste and caste-year fixed effects ( _φct_ and _ψsc_ ). The state-year fixed effects ( _πst_ ) take into account any differential trends across states in factors that affect fertility, son preference, and access to sex-selection technology. Like our savings specification, we also replace the main effect of _Dowrycst_ with caste _×_ state _×_ year fixed effects to estimate an even stricter specification that non-parametrically controls for everything that varies at the caste-state-year level and is correlated with fertility or sex-selection.

> 34In Table A.8, we re-estimate the effects on savings for families that only have one child. The sub-sample of one-child families offers a strict test for our story since the saving behavior of these families has not yet been affected by the differential likelihood of higher parity births or sex-selection of these births by firstborn sex (given that we flexibly control for duration since first birth effects). Despite the small sample size, we find that families that have only a girl child save significantly more than families that have only a boy child for given expected dowry. The higher magnitude of the savings effect in Table A.8 relative to Table 3 suggests that per capita savings fall as the number of children increases; however, given the small sample size, these are less reliable.

> 35NFHS interviewed women aged 15-49 at the time of the survey. We do not use the 1992-93 round of NFHS as it did not distinguish upper castes from other backward classes (OBCs)—which is essential for our identification strategy.

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