indicator has a positive and statistically significant effect on the change in MFN tariffs, as expected, since an increase in market power in a product provides an incentive to increase the MFN tariff of that product to benefit from the terms-of-trade improvement. In columns 2-9, we instrument for L.∆Prefik . The results in column 2 are similar to those of column 1. As before, when we redefine our dependent and main explanatory variables to include AD, in column 3, we find no effect of preferential liberalization on protection against the ROW. The market power variable is now not significant.

For the customs union period (columns 4-6) and the whole period (columns 7-9), we use an indicator for the change in market power, as defined above, rather than its level. We do not find evidence that our measure for the change in market power arising from the formation of the customs union has affected import protection against non-MERCOSUR countries. Equally important, our previous results regarding our main explanatory variable, L.∆Prefik , already discussed in section 4, remain qualitatively and quantitatively similar.

Panel B shows the results from analogous specifications for the case of Brazil. Again, we do not find any systematic evidence of an effect of our market power measures, and the results regarding the effect of preferential liberalization on external protection from table 5 remain unaltered.<sup>44</sup> The elasticity of export supply in an industry can also be endogenous to the level of import protection that exists in that industry, and there is measurement error in the estimated elasticities; therefore, we also tried instrumenting for the market power variable. We used the average market power in the other MERCOSUR countries in the same good as instrument.<sup>45</sup> Again, we did not find any systematic 44 As robustness, we also tried using the level of the market power variable instead of the indicator. We tried both the level and the log of the inverse export supply elasticity (and their changes due to the customs union), but overall the variable was less significant or had the wrong sign in some cases (and we also tried winsorizing the elasticities by setting the extreme values at the values at the 5th and 95th percentiles of the distribution). However, our results regarding the effect of preferential liberalization and the effects of including TTBs in the estimations remain robust. The results are also robust to working with the elasticities at the 4-digit HS level from the estimates from Nicita, Olarreaga and Silva (2015), concorded similarly, but overall the market power variable was less significant in some cases. Furthermore, we also tried adding the interaction of the market power variable (in levels, logs or the indicator) with the preferential liberalization variable, but the interaction was not significant.

45 Since we cannot use an IV-ordered probit when the endogenous regressor is not continuous, we do not employ the medium-high inverse export supply elasticity indicator. Instead, we use the level of the inverse export supply elasticity, and we set the top 5 percent of the elasticity data to the value at the 95<sup>th</sup> percentile of the distribution, and the bottom 5 percent to the value at the 5<sup>th</sup> percentile of the distribution, to deal with extreme values of the estimated elasticities.

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