to 47 months prior to the survey, while in NFHS-2 the same extends only to 1060 children born up to 36 months prior. NFHS-3 is the most comprehensive, covering 1256 children born up to 60 months prior to the survey date. We conduct all our analysis restricting our sample to those born up to 36 months before to be consistent with the most restrictive age limit, NFHS-2. This is reflected in the summary statistics for child age in each sample (Table 2b). The benefit of restricting the sample is to minimize possible recall or survival bias, at the cost of sample size.

We examine impacts on early-childhood mortality, child health status, and parental investments in child health, using this sub-sample of children aged up to 36 months. We match all observations to mother’s birth history data and to household-level data in order to compute eligibility for each individual child.

To examine the program effects here, we adopt the same strategy, computing the pre- and post program differences between eligible and non-eligible children. Because we have an additional source of variation by gender (since only girls may receive the benefits), we can further account for underlying time-varying trends that differentially affect eligible and non-eligible children, using the pre- and post program differences between eligible and non-eligible boys as a further comparison group. This leads to a triple-differenced linear specification for outcome _Yi_ for any child _i,_ , where where _I_<sup>_c_</sup> _i_<sup>is an indicator of an individual child’s eligibility excluding the gender criterion</sup> (1 if the child is the first, second or third child and their household is BPL or SC/OBC and 0 otherwise). _Fi_ is an indicator for being female, and _Tt_ are dummies for each survey wave. This specification captures differential trends over time for female children (  2f and  3f ) and eligible children (  2I and  3I ) as well as any underlying baseline differences specific to female children in eligible households (  f). Again, _X_ is a vector of household and individual level controls, this time including BPL status, SC/OBC status, gender and birth order.

First, we look at the available health status outcomes including children’s mortality and height-and-weight-for-age scores. We examine children’s mortality (reported death) at 0, 6, 12 and 24 months since birth<sup>10</sup> . Next, for the sample of all living children, we look at height-for-age and weight-for-age z-scores, as well as the incidence of stunting (a heightfor-age score under -2).

For the subsample of living children aged up to 36 months, we also examine immunization. However, a large fraction of the children in the sample lack a formal

> 10 The distribution of reported age at death in the NFHS has some marked irregularities (see Appendix Table A). In particular, while a spike in number of deaths recorded at birth (0 months) is plausible, the distribution also suggests that 6, 12, 18, 24 and 36 months are disproportionately represented salient focal points where bunching occurs, while the latter is not applicable to the NFHS-2 sample. We therefore use broad classifications of mortality to somewhat account for this.

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