fertility. Only foreign raised women (“abroad”) have lower chances of having their first birth while teenagers.

# 4.4 Robustness Checks

**_Alternative weighting scheme -_** When using all four rounds of the Kenyan DHS, we have observations from 4 rounds<sup>17</sup> from 1954 to 1969, observations from 3 rounds<sup>18</sup> for 1970 to 1973, observations from 2 rounds<sup>19</sup> for 1974 to 1978 and observations from the 2003 DHS for years 1979 and 1980. As a result, we are giving more weight to the oldest cohorts. If there is no difference in the effect of compulsory schooling laws on teenage childbearing over time, the results should be the same whether we weight each cohort equally or not. This, however, is testable. In Table 6, we present the results when we weight each cohort equally (thereby weighting each observation by the inverse of the weighted number of individuals in that cohort in our sample). While the results are consistent with those in Table 5, it does seem that, when less weight is placed on the earlier periods, the Reform is slightly more effective (all point estimates are slightly higher but not significantly different).

**_Cohorts around the threshold -_** So far, we used all cohorts from 1955 to 1980 in our regressions. However, we also saw in Figure 5 that some students already spent more time in school prior to the official implementation of the Reform. We explained this as the effect of grade repetition (girls born before 1971 who repeated a grade were subject to the Reform), later school start or pause between primary and secondary school. To avoid the problematic cohorts (1968 to 1971), we isolated the cohorts from 1955 to 1968 and 1971 to 1980 as comparison and treatment groups. Table 7 shows the results of equation (1) in that setting: girls who completed at least primary school are 12.9 percentage points less likely to become pregnant when they have one more year of education. Girls who have completed at least secondary school are 17.2 percentage points less likely to become pregnant. These are to compare with 18.8 and 19.3 percentage points. Although we would expect the new point estimates to be higher, those we obtained are not statistically significantly different from the point estimates of Table 5.

In all regressions, we included polynomial time-trends to control for the decrease over time of fertility and the increase in age at first birth. However, the youngest and oldest cohorts may not be relevant for the analysis: as we use a RDD, we hope to identify the effect of education around the threshold, i.e. on cohorts born close to 1971. Table 8 shows the same specification (equation (1)) for cohorts born between (1965-1968) and (1971-1974). Isolating cohorts around the Reform threshold, we get a first-stage point estimate of the impact of the Reform on education that is close to 1 for girls with at least a primary education degree (1.2). It magnitude (slightly higher than one) can be explained by the increase in years spent at school of almost 2 years of the population with at least a secondary degree: since the 1985 Reform was hastily put into place, universities had too little time to adapt to the increased number of incoming students, which may reflect why some repeated one year. The results of the second-stage are still positive and highly significant: girls who completed at least primary (respectively secondary) school are 6.6 (10) percentage points less likely to become pregnant when they have one more year of education. The smaller magnitude of the coefficients can be explained by the choice of a smaller window of observations and may indicate that the original specification does not capture well enough timetrends.

> 17 KDHS-89, KDHS-93, KDHS-98 and KDHS-03. 18 KDHS-93, KDHS-98 and KDHS-03. 19

> KDHS-98 and KDHS-03.

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