changes in their magnitudes. Therefore, both the main and auxiliary predictions of the model are veri fi ed and the magnitude of the stumbling block effect is similar, as noted in table 5.

As discussed in section 4.1, the data to construct an exact measure to capture the MFN externality is not available. Although we don’t think this seriously biases our results, as previously discussed, we create a proxy for this effect. For each HS-8-digit product, we calculated the share of the top 5 exporters in total exports to the EU in 1994 and 1989—the earliest year when the 8-digit hs data is reported. If a decrease in this share in a given product re fl ects a longer term decrease over the period between 1978 and 1994, then we can use it as a proxy. The model hence predicts a negative coefficient on this variable: products where the share increases have bigger decreases in tariffs.<sup>48</sup> When we include our proxy for the (inverse of) MFN externality we fi nd that it has the expected negative sign but it is insigni fi cant. This could either be because the effect is partially captured by the reciprocity variable, because the proxy is an imperfect one or MFN externalities were not signi fi cant. More importantly, for our purposes both the main and auxiliary predictions of the model are again con fi rmed and the growth effect is also statistically identical to the one in the baseline speci fi cation.

The theoretical model that guides our empirical estimation is parsimonious and therefore it is possible that we have not accounted for some determinants of the EU’s tariff changes. We do not expect that such omitted variables will bias our estimates because, as we argue above, even if they are correlated with the included regressors we instrument, test, and con fi rm the orthogonality of the excluded instruments relative to the error term. Nevertheless, there is one variable that does not appear in the theoretical model that we want to address explicitly: the initial tariff rate. The average MFN tariff for PTA products in our sample is 7.57%, whereas it is 12.8% for the non-PTA products. Although in the UR no explicit formula was followed such that higher tariffs would be cut by more than the lower ones, it is certainly a possible outcome and may lead us to fi nd bigger cuts in the nonPTA products. When we add the initial tariff level, we fi nd that its coefficient is typically negative, so products with higher initial tariffs had slightly bigger cuts, but it is not always statistically signi fi cant. Moreover, the initial tariff does not affect the sign, magnitude or signi fi cance of the basic stumbling block effect. As shown in table 5 the relative growth effect evaluated at the average initial tariff is at least as large as the ones found in table 2. Therefore, the basic results in table 2 are not driven by differences in initial tariffs across products. Since the main results are not sensitive to the inclusion

> 48 At an extreme the EU is likely to offer larger reductions when it goes from importing a good from 10 different countries to just 1 because when the negotiation is done with one country, and it is the single exporter of the good, then there is no free-rider problem. That is there are no other exporters waiting to receive an MFN tariff reduction negotiated by some other country. The single exporter internalizes all the gains of the MFN reduction and will thus offer bigger tariff reductions to goods exported to the EU.

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