the absolute gains or losses for boys in firstborn-girl families (which would require examining the coefficients of _Firstborn girl_ x _Post_ indicators) because we include among controls _Firstborn girl_ x _Y ear_ fixed effects ( _ωtGj_ ). Replacing _ωtGj_ with _Firstborn girl_ x _Post_ indicators in specification (1) shows that, in firstborn-girl families, not only girls but boys too received greater immunization postultrasound (though the magnitude of gains is smaller for boys than for girls), while breastfeeding duration was significantly extended only for girls. These results are consistent with smaller sibship sizes in firstborn-girl families.<sup>43</sup>

# **6.2 Son-Biased Fertility Stopping**

For reasons detailed earlier, in Tables 5 and 6 we investigate if son-biased fertility stopping behavior changed subsequent to the availability of sex-selection technology using an OLS specification. The coefficient of _Firstborn Girl_ is positive and significant in both tables confirming that, pre-ultrasound, women whose first child was a girl were 3 p.p. more likely to give birth in a given year and had 0.189 more births than women with a firstborn son. Our estimates show that these differentials were significantly reduced once ultrasound technology became available. Estimates conditional upon mother fixed effects in column (6) of Table 5 are also similar. Column 3 in Table 6 implies that the pre-ultrasound gap in the number of births declined by 0.081 to 0.106, or by 40 to 50 percent. To put this in a wider perspective, the presence of HIV reduces the average number of births a woman has during her life-cycle by 0.15 (Shapira (2013)). Controlling for mother’s fertility preference and son preference, and their interactions with _Firstborn girl_ does not alter these effects in any way.<sup>44</sup> Table 7 shows that the relative fertility decline in firstborn-girl families is driven by a shift from having three or more children toward having two or fewer children.

Note that our fertility and EFM results remain similar if we restrict the NFHS sample to rural households from the states included in the REDS sample. Moreover, our results in Table 6 are robust to restricting the sample to women that have likely completed childbearing (i.e., women of age _≥_ 33 or _≥_ 37 at the time of survey), easing concerns about censoring, which may be more severe among firstborn-girl relative to firstborn-boy households in the post-period, driving our results. As a robustness check, we also estimate the effect on the hazard of birth using a Cox proportional hazard model. Table A.13 reports the hazard ratios from this exercise; our results remain the same.

# **6.3 Actual vs Desired Fertility**

Since sex-selective abortion allows parents to avoid unwanted children, we also test whether availability of ultrasound drove actual fertility closer to desired fertility. Our estimates in Table 6 imply large reductions in undesired fertility, which the first row shows was 0.115 births in firstborn-girl relative to firstborn-boy families. The coefficients of interest (those on the triple-interaction terms) are similar for actual fertility and for excess fertility (actual minus desired fertility). This again

> 43These results are available upon request. We find no significant change in boy survival alongside increases in girl survival in firstborn-girl families, however.

> 44Using a logit specification in Table 5 yields very similar results.

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