Do Exporters Respond to Both Tari s and Nominal Exchange ...€¦ · gin of exports can respond to...

Do Exporters Respond to Both Tariffs and Nominal Exchange

Rates? Evidence from Chinese Firm-Product Data∗

Zhe Chen† Yoshinori Kurokawa‡

July 25, 2019

Abstract

The Melitz (2003) model with exchange rates and sticky wages predicts that the extensive mar-

gin of exports can respond to both trade liberalization and nominal exchange rates, which might

not, however, be supported by data according to the well-known argument about temporary vs.

permanent shocks. Using Chinese firm-product data from 2000 to 2006, this paper empirically

tests this theoretical prediction and has three main findings. First, though smaller than that of

tariffs charged by trade partners, the coefficient of exporter numbers on nominal exchange rates

is also significant. In fact, as the beta coefficients and overall effects indicate, exchange rates

are much more important than believed. Second, reductions in tariffs and yuan depreciation

increased not only entry but also exit of exporters. Finally, the implied elasticity of aggregate

exports with respect to tariffs is 1.60 and that with respect to nominal exchange rates is 0.20,

which are in line with the estimates in the international real business cycle models.

JEL classification: F12, F14, F31

Keywords: Melitz model, exchange rates, tariffs, extensive margin of exports, trade elasticity,

China

∗We are very grateful to Bo-Young Choi, Gian Luca Clementi, Loretta Fung, Chang Hong, Sewon Hur,Isao Kamata, Tim Kehoe, Matthew Kidder, Bingjing Li, Bernabe Lopez-Martin, Ferdinando Monte, PeterMorrow, Hiroshi Mukunoki, Yang Shen, Nyeong Seon Son, Siqiang Yang, Haishan Yuan, and Yuan Yuan fortheir useful comments and suggestions. We also thank the participants at the Fall 2017 Midwest Macroeco-nomics Meetings, the Fall 2017 Midwest International Trade Conference, the 2018 AMES, the 77th AnnualMeeting of the JSIE, and the 40th Annual Conference of the KIEA. Chen gratefully acknowledges financialsupport from the National Natural Science Foundation of China (Grant Number 71803020), and Kurokawagratefully acknowledges financial support from the JSPS KAKENHI (Grant Number JP19K01644). Thispaper previously circulated under the title “Does Yuan Appreciation Weaken the Increase in Exporters dueto Trade Liberalization? Evidence from Chinese Firm-Product Data.”†School of International Trade and Economics, University of International Business and Economics. 1302

Boxue Building, 10 Huixin Dongjie, Chaoyang District, Beijing 100029, China. E-mail: [email protected].‡Faculty of Humanities and Social Sciences, University of Tsukuba. 1-1-1 Tennodai, Tsukuba, Ibaraki

305-8571, Japan. E-mail: [email protected].

1

1 Introduction

More and more Chinese firms have been entering foreign markets. In addition, China has

recently been conducting the simultaneous trade liberalization and managed floating currency

policy. As Figure 1 shows, China has faced significant changes in both tariffs, charged by

trade partners, and nominal exchange rates. These observations raise a question: Can the

number of exporters—the extensive margin (EM) of exports—respond to both tariffs and

nominal exchange rates?

Theoretically, the answer to this question is yes. The introduction of nominal exchange

rates and sticky wages into the Melitz (2003) model of heterogeneous firm trade predicts that

the EM of exports can respond to not only trade liberalization but also nominal exchange

rates (e.g., Rodrıguez-Lopez, 2011).1 Empirically, however, the answer might be no, according

to the well-known argument by Ruhl (2008) that firms do not change their exporting status

in response to temporary shocks (exchange rates or productivity) while some firms do change

it in response to permanent shocks (tariffs).

We empirically test the aforementioned prediction from the Melitz (2003) model.2 Specif-

ically, we use Chinese firm-product annual data from 2000 to 2006 with 170 trade partners

to test whether the EM of China’s exports responds to both changes in tariffs (charged by

trade partners) and changes in nominal exchange rates.

The benchmark regression results support the theoretical prediction. Though relatively

small compared to that on tariffs, the coefficient of China’s exporter numbers—the EM of

exports—on nominal exchange rates is also significant. In fact, as indicated by the beta

coefficients and overall effects, exchange rates are much more important than believed. This

pattern also holds for the export value/quantity per exporter—the intensive margin (IM) of

exports—at the product level.

When we estimate the coefficients of the entry and exit of China’s exporters, the regression

results suggest that reductions in tariffs, charged by its trade partners, and yuan depreciation

increased not only entry but also exit of exporters. This is in line with Feng et al. (2017), who

1As will be mentioned later, by developing a sticky-wage model of heterogeneous firm trade with exchangerates and endogenous markups, Rodrıguez-Lopez (2011) extends the discussion of the Melitz model to analyzethe response of the EM of trade to nominal exchange rates. Bergin and Lin (2008) also examine the responseto exchange rates, in particular, exchange rate regimes.

2Notably, as mentioned by Baggs et al. (2009), most empirical studies that have examined tariffchanges/trade liberalization include exchange rates as a control variable; however, the analysis is at theindustry level, not at the firm level.

2

also empirically document that the entry and exit of China’s exporters increased following

China’s World Trade Organization (WTO) accession in 2001. Finally, we construct another

measurement of the EM (IM) of exports that is based on export product variety from China’s

exporters. Changes in tariffs affect both the EM and IM of exports while changes in nominal

exchange rate only affect the IM of exports. Thus, while changes in nominal exchange rates

affect the EM of exports in terms of exporter numbers, they do not in terms of export variety

from exporters.

We further perform several robustness checks regarding trade modes (ordinary vs. pro-

cessing trade), export destinations (OECD vs. non-OECD countries; main trade partners),

balanced data, high frequency data (quarterly), real exchange rates, and sectoral and prod-

uct differences. We find that the coefficients on tariffs and exchange rates are significantly

different between processing and ordinary trade firms and that the coefficients are greater if

the export destinations are non-OECD countries. The benchmark results are robust to main

trade partners, balanced data, quarterly data, and real exchange rates. The coefficients of

the EM of exports on tariffs are significant in 8 sectors in China’s top 10 export sectors at

the HS2 level while the coefficients on exchange rates are significant in 9 sectors. We also

find that the coefficients on tariffs and nominal exchange rates are larger for differentiated

goods. Finally, we discuss both firm- and aggregate-level implications from our results. We

particularly want to emphasize that the implied elasticity of aggregate exports with respect

to tariffs is 1.60 and that with respect to nominal exchange rates is 0.20, which are in line

with the estimated (Armington) elasticities in the international real business cycle models.

Our study makes the following contributions to the literature on the EM of trade. Melitz

(2003) focuses on the response of the EM of trade to trade liberalization, and Rodrıguez-

Lopez (2011) focuses on the response of the EM of trade to nominal exchange rates. We

focus on the responses of the EM of trade to these two factors; in particular, we use Chinese

firm-product data to empirically investigate whether the EM of China’s exports responds to

both tariffs (charged by trade partners) and nominal exchange rates.

Of course, there are studies that have empirically investigated the responses of firms or

exporters to trade liberalization vs. exchange rates (e.g., Baggs et al., 2009 ; Fitzgerald and

Hallerz, 2017). They have, however, focused on real exchange rates while we shed light on

the significance of nominal exchange rates. Feenstra (1989) theoretically argues that the

responses of the import price to tariffs and exchange rates are symmetric, and empirically

3

verifies this argument using U.S. import prices of Japanese cars, trucks, and motorcycles.

Motivated by Feenstra (1989), Baggs et al. (2009) (BBF) empirically test, using Canadian

industry-firm annual data, if real exchange rates affect Canadian firm survival/exit, and

they find that the overall effect of the real exchange rate changes is comparable to that of

the Canada-U.S. Free Trade Agreement tariff changes, although the coefficient on the real

exchange rate changes is smaller than that on the tariff changes. Moreover, Fitzgerald and

Hallerz (2017) (FH) empirically test if Irish exporters respond symmetrically to tariffs and

real exchange rates using Irish firm-product annual data, and they find that the coefficient

on tariffs is larger than that on real exchange rates. Differences/similarities between our

paper, BBF, and FH are threefold. First, while FH and BBF both empirically test the

equivalence between tariffs and real exchange rates particularly for developed countries, our

paper compares the responses to tariffs and nominal exchange rates for the developing country

China. Second, our paper and FH both focus on the responses of exporters while BBF look

at all Canadian firms. Third, while our nominal results are in line with the real results by

BBF and FH in that the coefficient on exchange rates is smaller than that on tariffs, our

results based on beta coefficients and overall effects indicate that exchange rates are much

more important than believed. Therefore, these three papers are satisfactory complements.

Our paper also makes contributions to the literature on trade elasticity, in particular, EM

elasticity. Our results imply that the elasticity of exporter numbers with respect to tariffs

is 1.061; that with respect to nominal exchange rates 0.112; and that with respect to real

exchange rates 0.210. On the other hand, the estimated elasticity of exporter numbers with

respect to tariffs by Bas et al. (2017) is 3.83, which is much larger than ours. That with

respect to real exchange rates by Tang and Zhang (2012) is 0.17, which is close to ours. We

note that, while Bas et al. (2017) and Tang and Zhang (2012), respectively, focus on tariffs

and exchange rates, we compare tariffs vs. exchange rates. Our results also imply that the

elasticity of export variety with respect to tariffs is 0.0875; that with respect to nominal

exchange rates 0.00235; and that with respect to real exchange rates 0.0132. The estimated

elasticity of export variety with respect to tariffs by Feenstra and Kee (2007) is about 2,

which is much larger than ours. That with respect to real exchange rates by Colacelli (2010)

is 0.045, which is close to ours in that both are less than 0.1. To our knowledge, however, no

past studies have estimated the elasticity of exporter numbers or export variety with respect

to nominal exchange rates.

4

The rest of this paper is organized as follows. Section 2 provides theoretical motivation for

our empirical question. Section 3 documents the regression specification and data. Section 4

reports the regression results and robustness checks. Section 5 discusses firm- and aggregate-

level implications from our results. Section 6 concludes.

2 Theoretical Motivation

We introduce nominal exchange rates and sticky wages to the Melitz (2003) model of heteroge-

neous firm trade to guide our empirical analysis on the relationship between tariffs/nominal

exchange rates and the EM of exports. Our model is a slightly simplified version of the

Rodrıguez-Lopez (2011) model.3 Note that Melitz (2003) focuses on the response of the EM

of trade to trade liberalization; Rodrıguez-Lopez (2011) the response to nominal exchange

rates; and our paper the responses to both of these two factors.

There are two countries: Home and Foreign. There is a continuum of households in the

interval [0, 1], and each household provides labor to the production of differentiated goods in

each country. There are firms heterogeneous in productivity, and each firm produces a single

variety under monopolistic competition. Home and Foreign markets are segmented. As does

Rodrıguez-Lopez (2011), we introduce the exogenous nominal exchange rates defined as the

price of the Foreign currency in terms of the Home currency (e.g., China’s yuan per U.S.

dollar). We also assume that nominal wages are fixed in the Home currency.

The setup of the preferences and production and the derivation of the equilibrium ex-

porting cutoff productivity level for Home firms are standard and have been relegated to

Appendix A.1. Here, we only state hypotheses, which are derived from the theoretical model

and will be explored empirically.

(a) A decrease in tariffs, charged by partners, causes an increase in exporter numbers (the

EM of exports).

(b) A decrease in nominal exchange rates—appreciation—causes a decrease in exporter

numbers.

3While Rodrıguez-Lopez (2011) analyzes the Melitz (2003) model with nominal exchange rates, stickywages, and translog preferences (endogenous markup) in main text, he also analyzes the Melitz model withnominal exchange rates, sticky wages, and CES preferences (constant markup) in appendix and shows thatthe results are robust. We slightly simplify the latter CES preference (constant markup) model to focus onanalyzing the relationship between tariffs/nominal exchange rates and the EM of exports.

5

(c) In the more differentiated sector, the responses of exporter numbers to tariffs and

nominal exchange rates can be larger.

The intuition for the hypothesis (c) is that, for the same percent change in tariffs or nominal

exchange rates, if the elasticity of substitution is lower, that is, if the markup is higher, then

a change in profits is larger. It leads to a larger change in the exporting cutoff and thus a

larger change in the mass of exporters. Thus, in the more differentiated sector, the elasticity

of the mass of exporters with respect to tariffs/nominal exchange rates can be larger.

3 Regression Specification and Data

3.1 Regression Specification

3.1.1 Exporter Numbers and Export Value/Quantity per Exporter

Guided by the Melitz model with nominal exchange rates presented in the previous section,

we empirically investigate the responses of exporter numbers to tariffs and nominal exchange

rates at the product-country-year level. The benchmark regression is as follows:

ln(Eijt) =α0 + α1 ln(1 + Tariffijt) + α2 ln(NERjt) + Processingijt

+ ρi + θj + ηt + εijt(3.1)

Here, Eijt is the number of exporters of product i from China to country j at time t.4

Tariffijt is the tariff of product i charged by country j at time t. NERjt is the nominal

exchange rate, which is defined as China’s yuan per currency of country j at time t; thus, when

NER increases, China’s yuan depreciates. The trade mode may affect exporters’ behaviors;

thus, we control the processing trade dummy,5 Processingijt. When the product is exported

under the processing trade mode, the dummy is 1; otherwise, it is 0. Finally, we control the

product fixed effect ρi, country fixed effect θj and time fixed effect ηt.

If α1 is negative and α2 is positive, then the implication from the Melitz model with

nominal exchange rates is supported. When tariff decreases, the EM of exports increases.

4Here, the number of exporters within a product is corresponding to the number of exporters within anindustry in the theory. As will be shown in Section 3.2, we define a product at the HS6 level.

5Processing trade refers to importing all or part of the raw and auxiliary materials and re-exporting thefinished products after processing.

6

When China’s yuan appreciates (depreciates), the EM of exports decreases (increases).

We also empirically investigate the responses of export value/quantity per exporter—

the IM of exports—at the product-country-year level. In that case, Eijt is the export

value/quantity per exporter of product i from China to country j at time t.

3.1.2 Entry and Exit of Exporters

A change in exporter numbers could be caused by entry or/and exit of exporters. Thus,

we also investigate the responses of the entry and exit of exporters to tariffs and nominal

exchange rates, respectively. If the export value of a firm is 0 in year t − 1 but positive in

year t, then we consider this firm as a new exporter in year t. If the export value of a firm is

positive in year t−1 but 0 in year t, then we consider this firm as an exit exporter in year t.6

The benchmark regression is as follows:

ln(ENijt) =β0 + β1 ln(1 + Tariffijt) + β2 ln(NERjt) + Processingijt

+ ρi + θj + ηt + µijt(3.2)

Here, the regression specification is similar to equation (3.1). Now ENijt is the number of new

(exit) exporters of product i from China to country j at time t. When ENijt is the number of

new exporters, if α1 is negative and α2 is positive, then a decrease in tariffs causes additional

Chinese firms to enter foreign markets, whereas China’s yuan appreciation (depreciation)

causes less (more) Chinese firms to enter. When ENijt is the number of exit exporters, if

α1 is positive and α2 is negative, then a decrease in tariffs causes less Chinese firms to quit

foreign markets, whereas the appreciation (depreciation) of China’s yuan causes more (less)

Chinese firms to quit.

3.1.3 Export Product Variety

In this section, we construct another measurement of the EM (IM) of exports that is based on

export product variety from exporters. Applying the methodology in Hummels and Klenow

6We also use other definitions of new and exit exporters as robustness checks. If a firm never exportedin previous years and year t is the first year in which this firm exports, then we consider this firm as anew exporter in year t. If a firm quits foreign markets in year t and never returns to foreign markets, thenwe consider this firm as an exit exporter in year t. The results for these new definitions are robust. Note,however, that as shown in Section 3.2, our dataset is from 2000 to 2006 and thus the new definitions are notaccurate for an exporter that actually enters or exits foreign markets beyond this period.

7

(2005), we define the firm-country-product specific export variety at the HS4 level, which

captures the number of HS6 product varieties within each HS4 category that an exporter

exports to a destination country.7

The EM index for exports of HS4 category i from exporter k in China to country j in

year t is defined as follows:

EEVijkt =

∑h∈Iijkt xhkt∑h∈Iikt xhkt

, (3.3)

where Iikt is the set of all HS6 product varieties within HS4 category i that exporter k

exports to the world in year t; Iijkt is the set of HS6 product varieties within HS4 category i

in which exporter k has positive exports to country j in year t; and xhkt is the export value

of HS6 product variety h from exporter k to the world in year t.8 Thus, EEVijkt can be

thought of as a weighted count of varieties that exporter k exports to country j in year t,

relative to all varieties that exporter k exports to the world. If EEVijkt = 1, it means that

exporter k exports the whole set of product varieties within HS4 category i to country j.

If EEVijkt = 0, it means that exporter k does not export any product variety within HS4

category i to country j. When EEVijkt increases, the range of export varieties within HS4

category i from exporter k to country j expands. Thus, EEVijkt can be considered as a

measurement of the EM of exports of HS4 category i from exporter k to country j.

The IM index for exports of HS4 category i from exporter k in China to country j in year

t is defined as follows:

IEVijkt =

∑h∈Iijkt xhjkt∑h∈Iijkt xhkt

, (3.4)

where xhjkt is the export value of HS6 product variety h from exporter k to country j in year

t. Thus, IEVijkt equals exporter k’s exports to country j relative to its exports to the world in

year t, within HS4 category i. If IEVijkt = 1, it means that country j is the only destination

for HS4 category i from exporter k. When IEVijkt increases, the average export value of

HS4 category i from exporter k to country j increases. Thus, IEVijkt can be considered as a

measurement of the IM of exports of HS4 category i from exporter k to country j.

Finally, multiplying IEVijkt by EEVijkt produces the ratio of exporter k’s exports of HS4

7We do not use the number of HS8 product varieties within each HS6 category for two reasons. First, thecode of HS8 is not stable, which changes year by year. Second, the number of HS8 product varieties withineach HS6 category is small.

8While Hummels and Klenow (2005) set the world’s exports to country j as the base, we set exporter k’sexports to the world as the base.

8

category i to country j to its exports of the HS4 category i to the world.

We investigate the responses of these new measurements of the EM and IM of exports to

tariffs and nominal exchange rates, respectively. The benchmark regression is as follows:

ln(EVijkt) =γ0 + γ1 ln(1 + Tariffijt) + γ2 ln(NERjt) + Processingijkt

+ ρi + θj + τk + ηt + νijkt(3.5)

Here, EVijkt is either EEVijkt, the new EM index of exports, or IEVijkt, the new IM index

of exports. The independent variables are the same as equation (3.1). We also control the

exporter fixed effect τk. When EV is EEV , if γ1 is negative and γ2 is positive, then a decrease

in tariffs increases export variety from China’s exporters, whereas China’s yuan appreciation

(depreciation) decreases (increases) it. When EV is IEV , if γ1 is negative and γ2 is positive,

then a decrease in tariffs increases the export value per export variety from China’s exporters,

whereas China’s yuan appreciation (depreciation) decreases (increases) it.

3.2 Data

The nominal exchange rate and Consumer Price Index (CPI) data are from the International

Monetary Fund (IMF) at the annual level.9 The tariff data is from the United Nations

Conference on Trade and Development (UNCTAD)—Trade Analysis Information System

(TRAINS). The tariffs of each HS 6-digit level product charged by each country are at the

annual level. In this study, we use the effectively applied (AHS) tariff which is defined as the

lowest available tariff. In most cases, AHS tariffs are the same as most-favored nation (MFN)

tariffs. But in rare cases AHS tariffs are lower than MFN tariffs, which indicates that China

has a trade agreement with these products. China’s export data is from their customs agency

at the transaction level. We aggregate these transactions to the firm-product-country-year

level. Thus, we have the number of exporters and export value/quantity per exporter for

each HS 6-digit product. The HS code changed in 2002 at the HS6 level. In order to keep

the product category consistent, we firstly aggregate the product to the HS6 level and then

convert all products to HS1996 by using the concordance between HS2002 and HS1996. The

sample period is from 2000 to 2006.

Table 1 presents the data summary, which merges the tariff, exchange rate, and export

9We also present the results with the quarterly data in the robustness checks.

9

datasets. The first part of Table 1 demonstrates that the exporter and destination numbers

increase over time. From 2000 to 2006, the exporter numbers increase by 175 percent and

destination numbers increase by 43 percent. The product numbers are almost constant. The

second part of Table 1 demonstrates that the tariffs, charged by partners, decrease over

time.10 In 2000, the simple average tariffs are 9.6 percent, and in 2006 the tariffs are 8.1

percent. In addition, the simple average tariffs of the OECD11 countries remain stable, and

that of the non-OECD countries decrease by almost 30 percent from 2000 to 2006. Figure 2

shows the average evolution of tariffs charged by China’s selected trade partners. The selected

trade partners are China’s top 21 export destinations minus Hong Kong and Singapore. The

tariffs charged by Hong Kong and Singapore are almost zero, and thus we exclude these two

destinations. Germany, Netherlands, UK, Italy, France, Spain, and Belgium are members

of European Union during the sample period and charge the same tariffs, and thus we use

EU to represent them. The average log deviation of gross tariff from 200012 is stable for

most OECD countries except Mexico. However, this index decreased sharply for non-OECD

countries, in particular India, Thailand and Brazil. The pattern is similar to that of Table 1.

The third part of Table 1 and Figure 1 demonstrate the time trend of China’s nominal

effective exchange rates (NEERs). After China’s WTO accession in 2001, the NEERs de-

crease until 2005—depreciation. On July 21, 2005, the People’s Bank of China announced a

revaluation of the yuan and a reform of the exchange rate regime. After 2005, the NEERs

began to increase—appreciation—and do so until 2015.13

The last part of Table 1 demonstrates the trends of the EM and IM indexes of exports

based on export product variety from China’s exporters. From 2000 to 2006, the average

EM index of exports remains stable for both ordinary and processing trade. This result is

consistent with the pattern in the first part, which demonstrates that the number of products

is stable. During the same period, the average IM index of exports decreases by 6 percent for

ordinary trade and by 17 percent for processing trade. In addition, the EM index is larger

10Table 1 calculates the average of tariffs over the countries included in our merged dataset, while Figure1 calculates the average of tariffs over all the countries included in the original tariff dataset.

11Though Taiwan is not a member of OECD, the GDP per capital of Taiwan is similar to members ofOECD countries. Thus, we attribute Taiwan in the OECD group.

12We use coefficients on year dummies in country-by-country regression of ln(1+Tariff) on HS6 fixed effectsand year dummies.

13Here, an increase (decrease) in the NEERs means China’s yuan appreciation (depreciation), whereas anincrease (decrease) in the independent variable NER—defined as China’s yuan per a foreign currency—meansChina’s yuan depreciation (appreciation).

10

but the IM index is smaller for processing trade, which implies that the scope of product

varieties from China’s exporters is larger for processing trade.

3.2.1 Tariffs and Exchange Rates for Selected Trade Partners

In this section, we present more detailed data for tariffs and nominal exchange rates. Figure

3 shows the time trends of tariffs charged to China and nominal exchange rates for China’s

selected trade partners from 2000 to 2006.14 Here, the tariff data is at the annual level, and

the exchange rate data is at the monthly level.

The figures imply that both tariffs and nominal exchange rates might be crucial factors

that contribute to changes in the EM of China’s exports. In the case of the euro area, for

example, tariffs, charged by the euro area, decreased and the yuan depreciated from 2002 to

2004, implying that both tariff reductions and the yuan’s depreciation might have increased

China’s EM. In the case of Japan, the tariffs charged by Japan decreased and the yuan

first appreciated and then depreciated during the period 2001 to 2005, implying that, while

tariff reductions might have kept increasing it, the yuan’s appreciation and depreciation,

respectively, might have decreased and increased China’s EM.

4 Regression Results

4.1 Exporter Numbers and Export Value/Quantity per Exporter

Column 1 of Table 2 presents the coefficients of equation (3.1), that is, the coefficients of the

EM and IM of exports on tariffs and nominal exchange rates. According to the hypotheses

(a) and (b) in Section 2, a reduction in tariffs charged by trade partners would increase the

number of China’s exporters while the appreciation (depreciation) of China’s yuan would

decrease (increase) it. Our results in Column 1 verify them. If the tariffs decrease by 1

percent, the exporter numbers would increase by 1.061 percent. If China’s yuan appreciates

by 1 percent, the exporter numbers would decrease by 0.112 percent. The coefficient of the

EM of exports on tariffs is larger than that on nominal exchange rates. This result is similar

to the results of BBF and FH, though their focus is on real exchange rates.

14Here, Germany, Netherlands, Italy, France, Spain, and Belgium are labeled as Euro Area.

11

Table 2 also demonstrates that the reductions in tariffs, charged by trade partners, stim-

ulate the IM of exports from China and the appreciation (depreciation) of China’s yuan

decreases (increases) it. If the tariffs decrease by 1 percent, the export value per exporter

would increase by 0.541 percent. If China’s yuan appreciates by 1 percent, the export value

per exporter would decrease by 0.0868 percent. The results are similar for the export quan-

tity per exporter. If the tariffs decrease by 1 percent, the export quantity per exporter would

increase by 0.443 percent. If China’s yuan appreciates by 1 percent, the export quantity per

exporter would decrease by 0.0766 percent. Again, the coefficient on tariffs is larger than that

on nominal exchange rates. In addition, Table 2 demonstrates that the number of exporters

is smaller, but the export value/quantity per exporter is larger in the processing trade mode.

One concern is that the coefficients on tariffs and nominal exchange rates cannot directly

be interpreted as the relative importance of these two factors. Thus, Column 2 reports the

standardized beta coefficients based on the specification in Column 1.15 If the tariff decreases

by 1 standard deviation, the exporter numbers would increase by 0.0663 standard deviation.

If the nominal exchange rate appreciates by 1 standard deviation, the exporter numbers would

decrease by 0.226 standard deviation. Thus, the response of the EM of exports to nominal

exchange rates is even much larger than that to tariffs in terms of standard deviation. For

the IM of exports, we have the similar results.

In order to further interpret the relative importance of tariffs and nominal exchange rates,

we examine the overall effects of these two factors on the EM and IM of exports year by year.

Using the annual changes of tariffs and nominal exchange rates in Table 1, and coefficients in

Column 1, Table 3 presents the results. From 2000 to 2001, the average tariffs (charged by

partners) decreased and China’s yuan appreciated. The overall effects of tariffs and nominal

exchange rates are similar on exporter numbers. But the overall effect of nominal exchange

rates is larger than that of tariffs on the IM of exports. From 2002 to 2005, tariffs decreased

due to China’s entry into WTO and China’s yuan depreciated, both boosting the exports

from China. The overall effect of tariffs dominates that of nominal exchange rates, both on

the EM and IM of exports. From 2005 to 2006, the average tariffs are almost constant, while

China’s yuan appreciated by 2.3 percent. Thus, the overall effect of nominal exchange rates

dominates that of tariffs, both on the EM and IM of exports.

15The beta coefficient standardizes the OLS coefficient to capture the change in standard deviation unitsof the dependent variable in response to a one standard deviation increase in the independent variable.

12

4.2 Entry and Exit of Exporters

In this section, we investigate the exporter dynamics in greater details. Exporter numbers

depend on the entry of new exporters and the exit of incumbent exporters. Exporters’ entry

and exit decisions might have heterogeneous responses to the tariffs and nominal exchange

rates. Thus, we separately examine the responses. Column 1 of Table 4 shows that the

reductions in tariffs, charged by trade partners, increase not only entry but also exit of

exporters. If the tariffs decrease by 1 percent, the new exporter numbers would increase by

0.841 percent and exit exporter numbers would increase by 0.799 percent. The coefficient

of entry is larger than that of exit. Thus, the net response to the tariff reductions is an

increase in the exporter numbers. At first sight, the results for exit might appear odd. In

the Melitz model, a tariff reduction causes the entry of new exporters but does not cause

the exit of incumbent exporters. As Feng et al. (2017) argue, however, an extended Melitz

model can show that a tariff reduction increases both the entry and exit of exporters. This is

because an increase in exporters due to the tariff reduction could increase the competition in

foreign markets, forcing low productivity exporters to exit. Feng et al. (2017) also empirically

document that the entry and exit of China’s exporters increased following China’s WTO

accession in 2001.

The responses to exchange rates are similar to those to tariffs in that depreciation increases

not only entry but also exit of exporters, but the coefficients on exchange rates are smaller

than those on tariffs. If China’s yuan depreciates by 1 percent, new exporter numbers would

increase by 0.0617 percent and exit exporter numbers also would increase by 0.112 percent.

Again, to gauge the relative importance of the two factors, Column 2 reports the stan-

dardized beta coefficients based on the specification in Column 1. If the tariff decreases by 1

standard deviation, new exporter numbers would increase by 0.0557 standard deviation. If

the nominal exchange rate appreciates by 1 standard deviation, the new exporter numbers

would decrease by 0.134 standard deviation. Thus, the response of new exporter numbers to

nominal exchange rates is even much larger than that to tariffs in terms of standard deviation.

For exit exporter numbers, we have the similar results.

13

4.3 Export Product Variety

In this section, we investigate the responses to tariffs and exchange rates from the perspective

of export product variety from China’s exporters. Column 1 of Table 5 shows that the

reductions in tariffs, charged by trade partners, increase not only the EM index but also the

IM index of exports. However, the coefficient of the IM index is much larger. If the tariffs

decrease by 1 percent, the EM index would increase by 0.0875 percent and the IM index

would increase by 0.164 percent. The exchange rate fluctuation only affects the IM index

of exports. If China’s yuan depreciates by 1 percent, the IM index would increase by 0.126

percent. Column 2 reports the standardized beta coefficients based on the specification in

Column 1. If the tariff decreases by 1 standard deviation, the EM index would increase by

0.008 standard deviation and the IM index would increase by 0.006 standard deviation. If

the nominal exchange rate appreciates by 1 standard deviation, the IM index would decrease

by 0.153 standard deviation and the EM index change is not significant.

Thus, the results indicate that changes in tariffs affect the EM of exports in terms of both

the exporter numbers and the export variety from exporters. Changes in nominal exchange

rates, on the other hand, affect it in terms of exporter numbers but do not in terms of export

variety from exporters.

4.4 Robustness Checks

In this section, we examine whether trade modes, export destinations (OECD vs. non-OECD

countries; main trade partners), balanced data, data frequency, real exchange rates, and sec-

toral and product heterogeneity would affect the benchmark results.

Trade Modes

Processing trade refers to the activity of importing all or part of the raw and auxiliary materi-

als from abroad and re-exporting the finished products after processing or assembly by firms

within China. It is possible that the effect of an exchange rate appreciation on the export

price of assembled products is canceled by the effect on the import price of the materials for

assembly. Thus, the response to exchange rate fluctuations on processing trade is likely to

14

be different from that on ordinary trade.16 We thus divide trade into two groups: ordinary

and processing trade. The columns 3 and 4 of Table 2 show that the coefficients of exporter

numbers on tariffs and nominal exchange rates are similar for both trade modes. But the

coefficients of export value/quantity per exporter are larger for processing trade. Thus, in

terms of the IM of exports, firms doing the processing trades are more responsive to the tariff

and nominal exchange rate shocks. In addition, the columns 3 and 4 of Table 4 show that

the coefficients of entry and exit are similar for both trade modes. The columns 3 and 4 of

Table 5 show that the coefficients of the EM index based on export product variety are also

similar but those of the IM index are larger for processing trade. In sum, in terms of the EM

of exports, there are no significant differences between two trade modes. In terms of the IM

of exports, the coefficients for the processing trade are significantly larger.

OECD vs. Non-OECD Countries

After China became a WTO member in 2001, the tariffs charged by trade partners, especially

developing countries, decreased considerably; for example, the comparison among India and

developed countries (Figure 3). Table 1 also presents that the average tariffs charged by

OECD countries decreased by 3.3 percent from 2000 to 2006. During the same period, the

average tariffs charged by non-OECD countries decreased by 29.9 percent. Thus, we divide

the export destinations into two groups: OECD and non-OECD countries. Last two columns

in Tables 2, 4 and 5 show that the results are mixed. In terms of the EM of exports (exporter

numbers, new exporter numbers, and exit exporter numbers), the coefficients for non-OECD

countries are larger than those for OECD countries. In terms of the IM of exports (export

value/quantity per exporter), the coefficients on tariffs are similar between two groups, but

the coefficients on nominal exchange rates are larger for non-OECD countries.

Main Trade Partners

Our data covers about 170 destinations, many of which are small economies. The trade with

these small economies might be significantly affected by unobservable factors other than tar-

iffs and nominal exchange rates, such as political relationship. Thus, we restrict our sample

to the top 50 partners for robustness checks. The export value with these partners is about

16Marquez and Schindler (2007) argue that estimating the response to exchange rates should differentiatethe ordinary and processing trade products.

15

97.5 percent of China’s total export value. The results in Column 1 of Table 6 show that our

main findings are still robust.

Balanced Data

Our data is an unbalanced data at the product-destination level. As Table 1 has shown, the

product number is stable from 2000 to 2006, but the destination number expands quickly

during the same period. Thus, the new export destinations might bias our results. In order

to exclude this concern, we restrict our sample to the product-destination pairs that appear

in the whole period (2000-2006). The export value in the balanced data is about 87 per-

cent of that of the full sample. Column 2 of Table 6 shows the results. For the continuing

product-destination pairs, the coefficient on tariffs is smaller than that for the benchmark,

and the coefficient on nominal exchange rates is larger.

Quarterly Data

The fluctuations of the annual exchange rate data might be small. Naknoi (2015) investigates

the EM of exports using the quarterly U.S. bilateral trade data. Thus, we perform robustness

checks with the quarterly nominal exchange rate data. We conduct a stacked regression as

follows:

ln(Eijt) =α0 + α1 ln(1 + Tariffijt) +3∑

h=0

α2h ln(NERjt−h) + Processingijt

+ ρi + θj + ηt + εijt

(4.1)

Here,∑3

h=0 α2h measures the response to exchange rate fluctuations over the current and last

three quarters. Column 3 of Table 6 demonstrates that the results remain robust. Notably,

however, the coefficients for the quarterly data are smaller than those for the annual data.

This is probably because firms take time to react to the tariff and exchange rate changes.

Real Exchange Rates

The nominal and real exchange rates present very different movements in the United States

from 2000-2006 (Figure A1 in Appendix A.2). Motivated by this observation, we perform

robustness checks with the real exchange rates. Table 7 presents the results with the real

16

exchange rates. Notably, the results remain robust.

Sectoral Differences

The responses of the EM of exports to tariffs and exchange rates might be different across

sectors. In this section, we additionally examine the responses by China’s top 10 export

sectors at the HS2 level. These sectors account for 66 percent of China’s exports. These sec-

tors are Electrical equipment (21.9 percent), Mechanical appliances (18.1 percent), Apparel,

not knitted or crocheted (5.4 percent), Apparel, knitted or crocheted (4.5 percent), Optical

instruments (3.1 percent), Furniture (3 percent), Toys and sports requisites (2.8 percent),

Mineral products (2.5 percent), Footwear (2.5 percent), and Iron or steel (2.4 percent). The

numbers in parentheses are ratios of exports in each sector relative to all exports from China.

Table 8 shows that the coefficient on tariffs is significant in 8 sectors while that on exchange

rates is significant in 9 sectors. Thus, Chinese firms respond more against exchange rates

than against tariff changes.

Differentiated Goods

As implied by Chaney’s (2008) model and the hypothesis (c) in our model, the responses

of the EM of exports to tariffs and nominal exchange rates might also be different across

products.17 In the more differentiated sector, the responses of exporter numbers to tariffs and

nominal exchange rates can be larger. Here, following the classification from Rauch (1999),

we divide products into two groups: homogeneous and differentiated goods. Table 9 presents

the results. Diff Dummy is 1 if the product is a differentiated good; otherwise, it is 0. We find

that the coefficient on nominal exchange rates is larger for differentiated goods. We also find

the similar result for tariffs. These results are in line with some other empirical findings. Yi

and Biesebroeck (2012) showed, using data for China’s imports from 129 countries during the

period 2001–2006, that differentiated goods have the higher elasticity of the EM of exports

to China with respect to China’s tariffs than homogeneous goods. Colacelli (2010) found,

using 136 countries’ bilateral export data during the period 1981–1997, that differentiated

goods have the higher elasticity of the EM of exports with respect to exchange rates, though

real.

17As noted by Yi and Biesebroeck (2012), Chaney (2008) defines the EM of exports as how much newexporters export while almost all the empirical studies testing Chaney (2008) define it as the number offirms/products in exports.

17

5 Discussion

5.1 Firm-level Implications: Temporary vs. Permanent Shocks

Our firm-product level results have shown that the entry and exit of China’s exporters re-

sponded significantly to not only reductions in tariffs but also changes in exchange rates.

At first sight, this result seems to be inconsistent with the well-known argument by Ruhl

(2008) that firms do not change their exporting status in response to temporary shocks while

some firms do change it in response to permanent shocks. This is because if the changes in

exchange rates are considered as temporary shocks, then the coefficients of the entry and exit

of exporters on exchange rates would be insignificant.

However, it should be noted that, besides trade liberalization policy, China was conducting

significant currency policy during the period to be studied in our paper: undervalued currency

policy until 2005 and appreciation policy since 2005. This means that, in China from 2000

to 2006, it might be appropriate to consider changes in both tariffs and exchange rates as

non-temporary shocks. Then it is not odd that our results have shown significant coefficients

of the entry and exit of China’s exporters on both tariffs and exchange rates. In this sense,

our results can be consistent with Ruhl’s (2008) argument.

5.2 Aggregate-level Implications: The Trade Elasticity and the

EM Elasticity

In Section 4, we have demonstrated that China’s exports significantly respond to both tariffs

and exchange rates. This is essentially a problem of the so-called elasticity of trade. Thus, it

is important to know whether the micro results in our paper are consistent with the response

of aggregate exports to these shocks in the existing studies.

We obtain the implied elasticity of aggregate exports by summing the coefficient of ex-

porter numbers and that of export value per exporter, which is averaged over HS6 products.

According to the results in Column 1 of Table 2, the elasticity of aggregate exports with

respect to tariffs is 1.60 and that with respect to nominal exchange rates is 0.20. According

to the results in Column 1 of Table 7, the elasticity of aggregate exports with respect to

real exchange rates is 0.34. According to Ruhl (2008), on the other hand, the estimated

(Armington) elasticities are 0.2-3.5 in the international real business cycle models and 4-15

18

in the applied general equilibrium models. Thus, all of our implied aggregate elasticities are

more close to the former estimates.

It should be, however, noted that our main interest has been particularly in the response

of the EM of exports, not the response of exports overall. Thus, it is also important to

know whether our results are consistent with the response of the EM of exports to tariffs

and exchange rates in the literature, although there are not many studies that estimated the

elasticity of the EM of exports with respect to tariffs and exchange rates. We summarize the

elasticities in our paper and other studies in Table 10.

We obtain the elasticity of the EM of exports by looking at the coefficient of exporter

numbers or export product variety. First, let us focus on exporter numbers as a measurement

of the EM of exports. According to the results in Column 1 of Table 2, the elasticity of

exporter numbers with respect to tariffs is 1.061, and that with respect to nominal exchange

rates is 0.112. According to the results in Column 1 of Table 5, the elasticity of exporter

numbers with respect to real exchange rates is 0.210. On the other hand, the estimated

elasticity of exporter numbers with respect to tariffs by Bas et al. (2017) is 3.83, which is

much larger than ours. That with respect to real exchange rates by Tang and Zhang (2012)

is 0.17, which is close to ours. Note that while our paper and Bas et al. (2017) are at the

product-destination-year level, Tang and Zhang (2012) are at the product-destination-month

level. To our knowledge, however, no past studies have estimated the elasticity of exporter

numbers with respect to nominal exchange rates.

Next, let us consider export variety as a measurement of the EM of exports. According

to the results in Column 1 of Table 8, the elasticity of export variety with respect to tariffs is

0.0875 and that with respect to nominal exchange rates is 0.00235. According to the results

in Column 1 of Table 9, that with respect to real exchange rates is 0.0132. According to the

literature, on the other hand, the estimated elasticity of export variety with respect to tariffs

by Feenstra and Kee (2007) is about 2, which is much larger than ours. That with respect

to real exchange rates by Colacelli (2010) is 0.045, which is larger than ours but close to

ours in that both are less than 0.1. Note that while our paper is at the firm-destination-year

level, Feenstra and Kee (2007) and Colacelli (2010) are at the industry-year level and the

country-destination-year level, respectively. Note also that, to our knowledge, no past studies

have estimated the elasticity of export variety with respect to nominal exchange rates.

19

6 Conclusion

We have empirically tested the implications from the Melitz model with the exchange rates.

Specifically, using China’s firm-product data from 2000 to 2006 with 170 trade partners, we

tested whether the EM of China’s exports respond to both trade liberalization and nominal

exchange rates. Based on regressions, we have three main empirical findings. First, though

relatively small compared to that of tariffs (charged by trade partners), the coefficient of

exporter numbers on nominal exchange rates is also significant. In fact, our results based on

the beta coefficients and overall effects indicate that exchange rates are much more important

than believed. Second, reductions in tariffs and yuan depreciation increased not only entry

but also exit of exporters. Finally, the implied elasticity of aggregate exports with respect to

tariffs is 1.60 and that with respect to nominal exchange rates is 0.20, which are in line with

the estimated (Armington) elasticities in the international real business cycle literature.

The results presented in this paper are valuable, particularly for empirical studies on trade

liberalization and the EM of trade. First, Kehoe and Ruhl (2013), for example, focused on

increases in the EM of trade after trade liberalization or structural changes. Our results,

however, indicate that the observed increases in the EM of trade could be caused by not only

changes in tariffs but also changes in nominal exchange rates. Thus, future studies, in par-

ticular, on China’s EM need to place more importance on exchange rates when investigating

the changes in the EM of trade using firm data. Second, our results provide an important

policy implication. China was a centrally-planned economy and adopted a U.S. dollar pegged

exchange rate policy for a long period. Previous studies paid little attention to the effect

of exchange rates on trade using Chinese data. China, however, allowed a managed float

of its currency in July 2005 and became the largest trader in the world in 2012. It is thus

important to examine the effect of exchange rates. In fact, our results indicate that foreign

currency policy could be an effective tool if Chinese government would like to boost exports.

20

References

Baggs, Jen, Eugene Beaulieu, and Loretta Fung (2009) ‘Firm survival, performance, and the

exchange rate.’ Canadian Journal of Economics 42(2), 393–421

Bas, Maria, Thierry Mayer, and Mathias Thoenig (2017) ‘From micro to macro: Demand,

supply, and heterogeneity in the trade elasticity.’ Journal of International Economics

108, 1–19

Bergin, Paul R., and Ching-Yi Lin (2008) ‘Exchange rate regimes and the extensive margin

of trade.’ NBER Working Paper 14126

Chaney, Thomas (2008) ‘Distorted gravity: The intensive and extensive margins of interna-

tional trade.’ American Economic Review 98(4), 1707–1721

Colacelli, Mariana (2010) ‘Intensive and extensive margins of exports and real exchange

rates.’ Working Paper

Feenstra, Robert C. (1989) ‘Symmetric pass-through of tariffs and exchange rates under

imperfect competition: An empirical test.’ Journal of International Economics 27(1), 25–

45

Feenstra, Robert C, and Hiau Looi Kee (2007) ‘Trade liberalization and export variety: A

comparison of Mexico and China.’ The World Economy 30(1), 5–21

Feng, Ling, Zhiyuan Li, and Deborah L. Swenson (2017) ‘Trade policy uncertainty and

exports: Evidence from China’s WTO accession.’ Journal of International Economics

106(1), 20–36

Fitzgerald, Doireann, and Stefanie Hallerz (2017) ‘Exporters and shocks.’ Federal Reserve

Bank of Minneapolis Staff Report 549

Guimaraes, Paulo, and Pedro Portugal (2010) ‘A simple feasible procedure to fit models with

high-dimensional fixed effects.’ Stata Journal 10(4), 628–649

Hummels, David, and Peter J Klenow (2005) ‘The variety and quality of a nation’s exports.’

American Economic Review 95(3), 704–723

21

Kehoe, Timothy, and Kim Ruhl (2013) ‘How important is the new goods margin in interna-

tional trade?’ Journal of Political Economy 121(2), 358–392

Marquez, Jaime, and John Schindler (2007) ‘Exchange-rate effects on China’s trade.’ Review

of International Economics 15(5), 837–853

Melitz, Marc J. (2003) ‘The impact of trade on aggregate industry productivity and intra-

industry reallocation.’ Econometrica 71(6), 1695–1725

Naknoi, Kanda (2015) ‘Exchange rate volatility and fluctuations in the extensive margin of

trade.’ Journal of Economic Dynamics & Control 52, 322–339

Rauch, James E (1999) ‘Networks versus markets in international trade.’ Journal of Inter-

national Economics 48(1), 7–35

Rodrıguez-Lopez, Jose A. (2011) ‘Prices and exchange rates: A theory of disconnect.’ Review

of Economic Studies 78, 1135–1177

Ruhl, Kim (2008) ‘The international elasticity puzzle.’ Working Paper, University of Texas

at Austin

Tang, Heiwai, and Yifan Zhang (2012) ‘Exchange rates and the margins of trade: Evidence

from Chinese exporters.’ CESifo Economic Studies 58(4), 671–702

Yi, Yingting, and Johannes Van Biesebroeck (2012) ‘The extensive margin of differentiated

goods and trade liberalization: Evidence from China.’ Working Paper

22

Table 1: Data Summary

Year

Variables 2000 2001 2002 2003 2004 2005 2006

Exporter, Product, and Destination

Exporter Numbers 59,709 65,505 74,882 90,705 114,768 138,881 163,923

Product Numbers 4,789 4,809 4,778 4,816 4,853 4,879 4,894

Destination Numbers 104 134 141 124 125 135 149

Tariffs Charged by Partners (%)

—All Countries

(mean) 9.60 9.54 9.97 8.86 7.89 8.11 8.10

(sd) 16.14 14.87 35.45 15.37 18.86 11.90 12.35

—OECD Countries

(mean) 4.91 5.01 5.12 4.98 3.93 4.63 4.75

(sd) 15.58 15.52 14.29 15.56 13.36 12.37 13.34

—Non-OECD Countries

(mean) 14.86 13.74 14.45 12.43 11.30 10.72 10.42

(sd) 15.12 12.90 46.76 14.28 21.99 10.83 11.04

China’s Nominal Effective Exchange Rate 93.18 98.35 97.92 91.88 87.68 87.26 89.23

Export Product Variety

—EM Index

Ordinary Trade 0.84 0.84 0.84 0.84 0.85 0.84 0.85

Processing Trade 0.92 0.92 0.92 0.93 0.93 0.93 0.93

—IM Index

Ordinary Trade 0.49 0.47 0.46 0.46 0.45 0.44 0.46

Processing Trade 0.36 0.35 0.34 0.33 0.32 0.32 0.30

Note: (a) Taiwan is not a member of OECD. However, the GDP per capital of Taiwan is similar to members of OECDcountries. Thus, we attribute Taiwan in the OECD group. (b) China’s nominal effective exchange rate index for 2010 is 100.

Source: Chinese Customs Export and Import Database, IMF and TRAINS.

23

Table 2: Nominal Exchange Rates, Tariffs, and Exports

(1) (2) (3) (4) (5) (6)

Full Sample Betas Ordinary Processing OECD Non-OECD

ln(Exporter Numbers)

ln(1+Tariff) -1.061*** -0.0663*** -1.041*** -1.220*** -0.341*** -0.662***

(0.0526) (0.00329) (0.0564) (0.0669) (0.0875) (0.0296)

ln(NER) 0.112*** 0.226*** 0.138*** 0.170*** 0.0521*** 0.168***

(0.00778) (0.0157) (0.00910) (0.0120) (0.0124) (0.00842)

Processing Trade -1.202*** -0.405*** -1.253*** -1.209***

(0.0196) (0.00661) (0.0197) (0.0229)

Observations 1,329,035 1,329,035 964,746 364,074 660,819 668,037

R-squared 0.577 0.577 0.633 0.606 0.667 0.531

ln(Export Value per Exporter)

ln(1+Tariff) -0.541*** -0.0223*** -0.410*** -1.057*** -0.337** -0.307***

(0.0627) (0.00259) (0.0597) (0.119) (0.137) (0.0491)

ln(NER) 0.0868*** 0.116*** 0.0631*** 0.256*** -0.0246 0.134***

(0.0138) (0.0183) (0.0141) (0.0286) (0.0245) (0.0151)

Processing Trade 0.591*** 0.132*** 0.654*** 0.426***

(0.0206) (0.00459) (0.0219) (0.0246)

Observations 1,329,035 1,329,035 964,746 364,074 660,819 668,037

R-squared 0.355 0.355 0.362 0.408 0.373 0.359

ln(Export Quantity per Exporter)

ln(1+Tariff) -0.443*** -0.0127*** -0.243*** -1.154*** -0.340** -0.330***

(0.0730) (0.00210) (0.0690) (0.130) (0.152) (0.0557)

ln(NER) 0.0766*** 0.0710*** 0.0609*** 0.259*** -0.0276 0.112***

(0.0149) (0.0138) (0.0158) (0.0305) (0.0268) (0.0165)


(0.0237) (0.00368) (0.0248) (0.0284)

Observations 1,329,035 1,329,035 964,746 364,074 660,819 668,037

R-squared 0.612 0.612 0.658 0.583 0.605 0.650

Product FE Yes Yes Yes Yes Yes Yes

Time FE Yes Yes Yes Yes Yes Yes

Country FE Yes Yes Yes Yes Yes Yes

Cluster by Product Yes Yes Yes Yes Yes Yes

Note: (a) Since the tariffs charged by Hong Kong are zero, we exclude Hong Kong from the sample. (b)Column 2 reports standardized beta coefficients from Column 1. (c) Taiwan is not a member of OECD.However, the GDP per capital of Taiwan is similar to members of OECD countries. Thus, we attribute Taiwanin the OECD group. (d) *** p < 0.01, ** p < 0.05, * p < 0.1.


24

Table 3: The Overall Effects of Tariffs and Nominal Exchange Rates Year by Year

Year

Variables 2000-2001 2001-2002 2002-2003 2003-2004 2004-2005 2005-2006

Exporter Numbers

Tariff 0.663 -4.576 13.292 11.616 -2.958 0.131

NER -0.620 0.049 0.690 0.510 0.020 -0.257

Export Value per Exporter

Tariff 0.338 -2.333 6.778 5.923 -1.508 0.067

NER -0.481 0.038 0.534 0.395 0.016 -0.200

Export Quantity per Exporter

Tariff 0.277 -1.911 5.550 4.850 -1.235 0.054

NER -0.424 0.034 0.472 0.349 0.014 -0.176


Table 4: Nominal Exchange Rates, Tariffs, and Exporter Dynamics

(1) (2) (3) (4) (5) (6)


ln(New Exporter Numbers)

ln(1+Tariff) -0.841*** -0.0557*** -0.924*** -0.837*** -0.218*** -0.512***

(0.0452) (0.00299) (0.0506) (0.0583) (0.0717) (0.0272)

ln(NER) 0.0617*** 0.134*** 0.0889*** 0.101*** 0.0758*** 0.0918***

(0.00820) (0.0178) (0.00926) (0.0130) (0.0150) (0.00891)


(0.0197) (0.00710) (0.0191) (0.0234)

Observations 1,082,359 1,082,359 825,716 256,341 523,419 558,736

R-squared 0.572 0.572 0.625 0.538 0.664 0.527

ln(Exit Exporter Numbers)

ln(1+Tariff) -0.799*** -0.0546*** -0.877*** -0.781*** -0.107 -0.531***

(0.0499) (0.00341) (0.0570) (0.0595) (0.0730) (0.0301)

ln(NER) 0.112*** 0.254*** 0.144*** 0.133*** 0.0731*** 0.147***

(0.00894) (0.0203) (0.0106) (0.0137) (0.0158) (0.00938)


(0.0187) (0.00712) (0.0185) (0.0223)

Observations 858,707 858,707 649,851 208,504 457,461 401,020

R-squared 0.567 0.567 0.616 0.542 0.649 0.518





Note: (a) Since the tariffs charged by Hong Kong are zero, we exclude Hong Kong from the sample. (b)Column 2 reports standardized beta coefficients from Column 1. (c) Taiwan is not a member of OECD.However, the GDP per capital of Taiwan is similar to members of OECD countries. Thus, we attribute Taiwanin the OECD group. (d) *** p < 0.01, ** p < 0.05, * p < 0.1.


25

Table 5: Nominal Exchange Rates, Tariffs, and Export Product Variety

(1) (2) (3) (4) (5) (6)


ln(EM Index)

ln(1+Tariff) -0.0875*** -0.00800*** -0.0835*** -0.102*** -0.0758** -0.0278*

(0.0253) (0.00231) (0.0265) (0.0268) (0.0320) (0.0163)

ln(NER) -0.00235 -0.00751 -0.00138 0.00245 0.0208*** 0.00943**

(0.00424) (0.0136) (0.00455) (0.00504) (0.00770) (0.00420)


(0.00631) (0.00261) (0.00662) (0.00758)

Observations 14,310,474 14,310,474 12,380,881 1,919,054 9,361,207 4,924,697

R-squared 0.130 0.130 0.128 0.153 0.130 0.150

ln(IM Index)

ln(1+Tariff) -0.164** -0.00574** -0.129** -0.592** -0.0380 -0.00245

(0.0700) (0.00245) (0.0558) (0.256) (0.137) (0.0413)

ln(NER) 0.126*** 0.153*** 0.120*** 0.238*** 0.167*** 0.0778***

(0.0138) (0.0168) (0.0138) (0.0285) (0.0240) (0.0134)


(0.0406) (0.00643) (0.0388) (0.0385)

Observations 14,310,474 14,310,474 12,380,881 1,919,054 9,361,207 4,924,697

R-squared 0.362 0.362 0.327 0.524 0.345 0.444




Exporter FE Yes Yes Yes Yes Yes Yes


Note: (a) Since the tariffs charged by Hong Kong are zero, we exclude Hong Kong from the sample. (b) Column2 reports standardized beta coefficients from Column 1. (c) Taiwan is not a member of OECD. However, theGDP per capital of Taiwan is similar to members of OECD countries. Thus, we attribute Taiwan in the OECDgroup. (d) *** p < 0.01, ** p < 0.05, * p < 0.1.


26

Table 6: Nominal Exchange Rates, Tariffs, and Exports:Robustness Checks

(1) (2) (3)

Main Trade Partners Full Balanced Quarterly

stacked


ln(1+Tariff) -1.068*** -0.881*** -0.895***

(0.0657) (0.0851) (0.0566)

ln(NER) 0.0996*** 0.221*** 0.0916***

(0.00953) (0.0132) (116.56)

Processing Trade -1.302*** -1.284*** -0.845***

(0.0191) (0.0249) (0.0184)

Observations 985,575 552,956 3,633932

R-squared 0.607 0.613 0.523


ln(1+Tariff) -0.655*** -0.304*** -0.365***

(0.0782) (0.0904) (0.0607)

ln(NER) 0.0883*** 0.155*** 0.0384***

(0.0165) (0.0180) (7.8)

Processing Trade 0.651*** 0.925*** 0.568***

(0.0202) (0.0245) (0.0201)

Observations 985,575 552,956 3,633,932

R-squared 0.337 0.467 0.329


ln(1+Tariff) -0.600*** -0.342*** -0.266***

(0.0882) (0.106) (0.0727)

ln(NER) 0.0860*** 0.0992*** 0.0354**

(0.0183) (0.0193) (5.23)

Processing Trade 0.478*** 0.776*** 0.432***

(0.0231) (0.0271) (0.0236)

Observations 985,575 552,956 3,633,932

R-squared 0.613 0.667 0.599

Product FE Yes Yes Yes

Time FE Yes Yes Yes

Country FE Yes Yes Yes

Cluster by Product Yes Yes Yes

Note: (a) Since the tariffs charged by Hong Kong are zero, we excludeHong Kong from the sample. (b) The export value of the top 50 partnersis about 97.5 percent of China’s total export value. (c) The stacked re-gression in Column 3 measures the responses to exchange rate fluctuationsover the current and last three quarters. (d) *** p < 0.01, ** p < 0.05, * p < 0.1.


27

Table 7: Real Exchange Rates, Tariffs, and Exports

(1) (2) (3)

ln(Exporter Numbers) ln(Export Value per Exporter) ln(Export Quantity per Exporter)

ln(1+Tariff) -1.068*** -0.530*** -0.447***

(0.0535) (0.0638) (0.0743)

ln(RER) 0.210*** 0.131*** 0.140***

(0.00943) (0.0165) (0.0184)

Processing Trade -1.198*** 0.606*** 0.448***

(0.0195) (0.0205) (0.0236)

Observations 1,278,015 1,278,015 1,278,015

R-squared 0.577 0.354 0.611

Product FE Yes Yes Yes

Time FE Yes Yes Yes

Country FE Yes Yes Yes

Cluster by Product Yes Yes Yes

(1) (2)

ln(New Exporter Numbers) ln(Exit Exporter Numbers)

ln(1+Tariff) -0.850*** -0.794***

(0.0460) (0.0506)

ln(RER) 0.210*** 0.113***

(0.00959) (0.00970)

Processing Trade -1.274*** -1.194***

(0.0196) (0.0186)

Observations 1,039,128 829,596

R-squared 0.573 0.567

Product FE Yes Yes

Time FE Yes Yes

Country FE Yes Yes

Cluster by Product Yes Yes

(1) (2)

ln(EM Index) ln(IM Index)

ln(1+Tariff) -0.0860*** -0.154**

(0.0251) (0.0713)

ln(RER) 0.0132*** 0.160***

(0.00510) (0.0166)

Processing Trade 0.123*** -0.314***

(0.00633) (0.0406)

Observations 13,871,508 13,871,508

R-squared 0.130 0.361

Product FE Yes Yes

Time FE Yes Yes

Country FE Yes Yes

Exporter FE Yes Yes

Cluster by Product Yes Yes

Note: (a) Since the tariffs charged by Hong Kong are zero, we exclude Hong Kong from the sample. (b) *** p < 0.01, ** p <0.05, * p < 0.1.


28

Table 8: Exchange Rates, Tariffs, and Exports by Sectors

(1) (2) (3) (4) (5) (6)

Nominal Exchange Rate Real Exchange Rate

Numbers Value Quantity Numbers Value Quantity

Electrical equipment Tariff -1.054*** -1.189*** -1.162*** -1.022*** -1.148*** -1.127***

(0.153) (0.222) (0.273) (0.156) (0.226) (0.276)

Exchange Rate 0.172*** 0.194*** 0.172*** 0.229*** 0.339*** 0.347***

(0.0201) (0.0457) (0.0495) (0.0249) (0.0507) (0.0613)

Mechanical appliances Tariff -0.831*** -0.484* -0.153 -0.801*** -0.470* -0.150

(0.138) (0.248) (0.307) (0.141) (0.250) (0.310)

Exchange Rate 0.157*** 0.142*** 0.211*** 0.216*** 0.171*** 0.303***

(0.0232) (0.0454) (0.0500) (0.0251) (0.0489) (0.0578)

Apparel, not knitted or crocheted Tariff 0.743*** 0.586*** 0.938*** 0.768*** 0.576*** 0.930***

(0.214) (0.197) (0.265) (0.211) (0.197) (0.260)

Exchange Rate 0.526*** 0.303*** 0.359*** 0.590*** 0.326*** 0.404***

(0.0349) (0.0625) (0.0680) (0.0442) (0.0684) (0.0757)

Apparel, knitted or crocheted Tariff -0.476** -0.0448 0.387 -0.418** -0.0347 0.391

(0.199) (0.313) (0.363) (0.202) (0.310) (0.361)

Exchange Rate 0.310*** 0.0668 0.0897 0.315*** 0.113 0.0663

(0.0394) (0.0552) (0.0708) (0.0492) (0.0774) (0.0896)

Optical instruments Tariff -0.576* -1.187*** -1.522*** -0.556 -1.076** -1.498***

(0.332) (0.413) (0.437) (0.330) (0.412) (0.425)

Exchange Rate 0.407*** 0.326*** 0.234*** 0.729*** 0.621*** 0.571***

(0.0451) (0.0920) (0.0858) (0.0684) (0.122) (0.106)

Furniture Tariff -0.00981 0.00599 0.199 -0.0647 -0.0427 0.130

(0.429) (0.560) (0.587) (0.429) (0.561) (0.593)

Exchange Rate 0.349*** 0.209*** 0.109* 0.507*** 0.291*** 0.223**

(0.0415) (0.0566) (0.0648) (0.0577) (0.0739) (0.0828)

Toys and sports requisites Tariff -0.334 -0.0972 -0.377 -0.316 -0.0625 -0.386

(0.449) (0.429) (0.407) (0.426) (0.408) (0.388)

Exchange Rate 0.170** 0.114 0.0294 0.277*** 0.135 0.0162

(0.0710) (0.0784) (0.0878) (0.0895) (0.0939) (0.0969)

Mineral products Tariff -0.793*** -0.0590 -0.0270 -0.845*** -0.0768 -0.0920

(0.210) (0.323) (0.347) (0.206) (0.322) (0.345)

Exchange Rate 0.209*** 0.272*** 0.197*** 0.335*** 0.234*** 0.214***

(0.0310) (0.0617) (0.0660) (0.0461) (0.0744) (0.0777)

Footwear Tariff -0.601* -0.0748 0.0485 -0.624** -0.0542 0.0580

(0.308) (0.387) (0.422) (0.308) (0.385) (0.416)

Exchange Rate 0.0737** 0.148** 0.182** 0.179*** 0.222** 0.359***

(0.0296) (0.0598) (0.0757) (0.0360) (0.0869) (0.0952)

Iron or steel Tariff -1.938*** -1.943*** -2.147*** -1.998*** -2.036*** -2.278***

(0.211) (0.371) (0.386) (0.218) (0.392) (0.404)

Exchange Rate 0.0690 0.0642 0.0110 0.241*** 0.142* 0.0833

(0.0431) (0.0754) (0.0726) (0.0536) (0.0854) (0.0881)

Note: (a) Since the tariffs charged by Hong Kong are zero, we exclude Hong Kong from the sample. (b) *** p < 0.01, ** p <0.05, * p < 0.1.


29

Table 9: Exchange Rates, Tariffs, and Differentiated Products

(1) (2) (3) (4)

Nominal Exchange Rate Real Exchange Rate


ln(1+Tariff) -1.135*** 0.0483 -1.144*** 0.0730

(0.0542) (0.247) (0.0552) (0.100)

ln(1+Tariff) × Diff Dummy -1.352*** -1.380***

(0.290) (0.111)

ln(Exchange Rate) 0.113*** 0.104*** 0.204*** 0.164***

(0.00809) (0.0334) (0.00978) (0.00942)

ln(Exchange Rate) × Diff Dummy 0.0475*** 0.0496***

(0.00958) (0.00251)


(0.0202) (0.0331) (0.0201) (0.0201)

Observations 1,247,334 1,245,966 1,199,222 1,199,222

R-squared 0.577 0.600 0.577 0.580


ln(1+Tariff) -0.583*** 0.290** -0.576*** 0.298**

(0.0673) (0.129) (0.0685) (0.130)


(0.138) (0.139)

ln(Exchange Rate) 0.0879*** 0.0603*** 0.131*** 0.101***

(0.0142) (0.0145) (0.0170) (0.0172)


(0.00389) (0.00403)


(0.0212) (0.0212) (0.0210) (0.0210)

Observations 1,247,334 1,247,334 1,199,222 1,199,222

R-squared 0.354 0.355 0.352 0.353


ln(1+Tariff) -0.478*** 0.304** -0.486*** 0.299**

(0.0788) (0.138) (0.0803) (0.139)


(0.150) (0.151)

ln(Exchange Rate) 0.0810*** 0.0367** 0.141*** 0.0929***

(0.0154) (0.0158) (0.0190) (0.0192)


(0.00453) (0.00469)


(0.0245) (0.0245) (0.0244) (0.0244)

Observations 1,247,334 1,247,334 1,199,222 1,199,222

R-squared 0.618 0.618 0.617 0.618

Product FE Yes Yes Yes Yes

Country FE Yes Yes Yes Yes

Time FE Yes Yes Yes Yes

Cluster by Product Yes Yes Yes Yes

Note: (a) Since the tariffs charged by Hong Kong are zero, we exclude Hong Kongfrom the sample. (b) Diff Dummy is 1 if the product is a differentiated good according tothe classification from Rauch (1999); otherwise, it is 0. (c) *** p < 0.01, ** p < 0.05, * p < 0.1.


30

Table 10: Elasticities in Selected Studies

Paper Country Tariff elasticity Exchange rate elasticity

NER RER

Exporter numbers

Ours China 1.061 0.112 0.210

Bas et al. (2017) China & France 3.83

Tang and Zhang (2012) China 0.17

Export variety

Ours China 0.0875 0.00235 0.0132

Feenstra and Kee (2007) Mexico 2.049

Colacelli (2010) 136 countries 0.045

Note: (a) For exporter numbers, our paper and Bas et al. (2017) are at the product-destination-year level,and Tang and Zhang (2012) are at the product-destination-month level. (b) For export variety, our paperis at the firm-destination-year level; Feenstra and Kee (2007) are at the industry-year level; and Colacelli(2010) is at the country-destination-year level.

31

Figure 1: Average Tariffs Charged by Trade Partners and China’s Nominal Effective Ex-change Rates

9010

011

012

013

0N

EER

89

1011

12Ta

riff

2000 2005 2010 2015Year

Tariff NEER

Note: An increase (decrease) in nominal effective exchange rates means China’s yuan appreciation(depreciation).

Source: IMF and TRAINS.

32

Figure 2: Average Evolution of Tariffs Charged by Selected Trade Partners

-.06

-.04

-.02

0.0

2

avg

log

devi

atio

n of

gro

ss ta

riff f

rom

200

0

2000 2002 2004 2006Year

USA EU JPN

KOR TWN CAN

AUS MEX

OECD

-.2-.1

5-.1

-.05

0

avg

log

devi

atio

n of

gro

ss ta

riff f

rom

200

0

2000 2002 2004 2006Year

IND BRA

THA IDN

MYS

non-OECD

Note: (a) This figure shows coefficients on year dummies in country-by-country regression of ln(1+Tariff)on HS6 fixed effects and year dummies. (b) Taiwan is not a member of OECD. However, the GDP percapital of Taiwan is similar to that of members of OECD countries. Thus, we attribute Taiwan to theOECD group.

Source: TRAINS.

33

Figure 3: Tariffs Charged to China and Nominal Exchange Rates for Selected Trade Partners

7.8

7.9

88.

18.

28.

3Yu

an/U

SD

3.5

3.6

3.7

3.8

3.9

4Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

United States

78

910

11Yu

an/E

uro

4.2

4.4

4.6

4.8

5Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Euro Area

.06

.065

.07

.075

.08

Yuan

/Yen

1.4

1.6

1.8

2Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Japan

.006

.006

5.0

07.0

075

.008

.008

5Yu

an/W

on

9.5

1010

.511

11.5

Tarif

f

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

South Korea

1213

1415

16Yu

an/P

ound

4.2

4.4

4.6

4.8

5Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

United Kingdoms

.24

.25

.26

.27

Yuan

/TW

D

4.5

55.

56

6.5

Tarif

f

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Taiwan

55.

56

6.5

77.

5Yu

an/C

AD

2.6

2.8

33.

23.

4Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Canada

44.

55

5.5

66.

5Yu

an/A

UD

3.5

44.

55

5.5

Tarif

f

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Australia

2.14

2.16

2.18

2.2

2.22

Yuan

/MYR

7.8

88.

28.

4Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Malaysia

.17

.175

.18

.185

.19

Yuan

/INR

1520

2530

35Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

India

.000

7.0

008

.000

9.0

01.0

011

Yuan

/IDR

6.6

6.7

6.8

6.9

7Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Indonesia

.18

.19

.2.2

1.2

2Yu

an/T

HB

810

1214

16Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Thailand

.7.7

5.8

.85

.9Yu

an/M

XN

1214

1618

Tarif

f

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Mexico

23

45

Yuan

/BR

L

1314

1516

17Ta

riff

2000 2001 2002 2003 2004 2005 2006Year

tariff NER

Brazil

Note: A decrease (increase) in nominal exchange rates means China’s yuan appreciation (depreciation).

Source: IMF and TRAINS.

34

A Appendix

A.1 Theory

This appendix introduces nominal exchange rates and sticky wages to the Melitz (2003)

model of heterogeneous firm trade to guide our empirical analysis on the relationship between

tariffs/nominal exchange rates and the EM of exports. Our model is a slightly simplified

version of the Rodrıguez-Lopez (2011) model. There are two countries: Home and Foreign.

There is a continuum of households in the interval [0, 1], and each household provides labor

to the production of differentiated goods in each country. There are firms heterogeneous in

productivity, and each firm produces a single variety under monopolistic competition. Home

and Foreign markets are segmented.

First, we set up the preferences and production and show the equilibrium exporting cutoff

productivity level for Home firms. Then, we derive the proposition regarding the elasticity

of the exporting cutoff/the mass of exporters with respect to tariffs and that with respect to

nominal exchange rates.

Preferences

The representative Home household has the CES preferences for a continuum of differentiated

goods produced in Home and Foreign. The utility function is then given by

U =

(∫i∈∆′

qν−1ν

i

) νν−1

,

where qi is consumption of variety i, ν > 1 is the elasticity of substitution between varieties,

and ∆′= [0, N ] is the set of available varieties at Home.

Solving the utility maximization problem gives the demand for variety i

qi =(piP

)−νQ,

where pi is the price of variety i and P = (∫i∈∆′

p1−νi )

11−ν is the price of the composite good

Q. The total expenditure of this household is I = PQ. Since households are located in the

unit interval, the market demand and the representative household’s demand are equivalent.

35

Production

The production function of a Home firm with productivity φ is given by

y(φ) = φL,

where y(φ) is the output of a Home firm with productivity φ and L is the only factor of

production.18 As in the Melitz model, producers are heterogeneous in productivity, and each

producer knows its productivity, φ, only after entry. The productivity is Pareto distributed

in the interval [φmin,∞), and thus the cumulative distribution function for productivity is

G(φ) = 1− (φmin/φ)κ, where κ > 1 is a parameter of productivity dispersion (the higher κ,

the lower heterogeneity).

As does Rodrıguez-Lopez (2011), we assume that nominal wages are fixed at W in the

Home currency. Then the marginal cost of a Home firm with productivity φ is constant at

W/φ. Foreign production can analogously be defined. The production function of a Foreign

firm with productivity φ is just y∗(φ) = φL∗, where y∗ is the output of a Foreign firm with

productivity φ and L∗ is the Foreign labor. With fixed wages W ∗ in the Foreign currency,

its marginal cost is W ∗/φ.

There is the fixed cost of entry in terms of labor: fE and f ∗E for Home and Foreign firms,

respectively. There is also the fixed cost from selling in each market.19 The fixed cost from

selling in the Home market is fD for Home firms and f ∗X for Foreign firms; the fixed cost

from selling in the Foreign market is fX for Home firms and f ∗D for Foreign firms. Another

exporting cost is an iceberg cost. τ denotes the iceberg cost for Home firms so that a Home

exporter must ship τ > 1 units of the good in order for one unit to reach the Foreign market.

We interpret τ − 1 as tariffs imposed by Foreign. Similarly, τ ∗ denotes the iceberg cost for

Foreign firms, and we interpret τ ∗ − 1 as tariffs imposed by Home.

The profit maximizing domestic price in the Home currency for a Home firm with pro-

ductivity φ is given by

pD(φ) = (1 + µ)W

φ,

18In Rodrıguez-Lopez (2011), the production function of a Home firm with productivity φ is given byy(φ) = ZφL, where Z is an aggregate labor productivity factor. Because Z is not necessary for our purpose,we set Z = 1 for simplification.

19Without fixed costs in this CES model, it is always optimal for firms to produce a positive output foreach market.

36

where µ is the constant markup over marginal cost and equals

µ =1

ν − 1.

As does Rodrıguez-Lopez (2011), let ε be the exogenous nominal exchange rates defined as

the price of the Foreign currency in terms of the Home currency (e.g., China’s yuan per U.S.

dollar). Then the export price in the Foreign currency for a Home firm with productivity φ

is given by

pX(φ) = τ(1 + µ)W

εφ. (A.1)

Equilibrium Exporting Cutoff

Then, we can derive the equilibrium exporting cutoff level for Home firms:

φX = τ

[(fXf ∗D

)1νW

εW ∗

] νν−1

φ∗D, (A.2)

where φ∗D is the cutoff productivity level for Foreign firms selling in the Foreign market and

equals

φ∗D =f∗ 1κ

D

τF ∗ 1κ

[(ττ ∗)κF

κ−(ν−1)ν−1 − 1

τ ∗κFκ−(ν−1)ν−1 − 1

(ρε)κνν−1

(fXf∗D

)κ−(ν−1)ν−1

] 1κ

, (A.3)

where F = (fXf∗X)/(fDf

∗D), ρ = (F ∗/F )

ν−1κν (W ∗/W ), F ∗ = [κ− (ν − 1)]/(κφκmin)(δf ∗E +

f ∗D + f ∗X), F = [κ− (ν − 1)]/(κφκmin)(δfE + fD + fX) and δ is a proportion of existing firms

in each country to exit.20

Elasticity of the Exporting Cutoff and the Mass of Exporters

Then, we obtain the following proposition. Note that while the exporting cutoff part of the

proposition reproduces the results by Melitz (2003) and Rodrıguez-Lopez (2011), the mass

of exporters part was not explicitly derived by them.

Proposition:

20The equilibrium cutoff productivity levels can be derived using the zero-cutoff-profit conditions and thefree-entry conditions.

37

(a) The elasticity of the exporting cutoff with respect to tariffs for Home firms is:

∂ lnφX∂ ln τ

= ζφX ,τ > 1,

and the elasticity of the mass of Home exporters with respect to tariffs is:

∂ lnNX

∂ ln τ= ζNX ,τ < −κ.

(b) The elasticity of the exporting cutoff with respect to nominal exchange rates for Home

firms is:∂ lnφX∂ ln ε

= ζφX,ε < −(1 + µ),

and the elasticity of the mass of Home exporters with respect to nominal exchange rates is:

∂ lnNX

∂ ln ε= ζNX ,ε > κ(1 + µ).

Proof:

(a) ζφX ,τ > 1:

We obtain the elasticity of the exporting cutoff with respect to tariffs τ by taking log of (A.2)

and differentiating it with respect to τ :

ζφX ,τ = 1 + ζφ∗D,τ ,

where ζφ∗D,τ is the elasticity of the domestic cutoff level (φ∗D) with respect to τ for Foreign

firms.

We next show that ζφ∗D,τ > 0. By taking log of (A.3) and differentiating it with respect

to τ , we obtain

ζφ∗D,τ = −1 +1

1− 1

ττ∗κFκ−(ν−1)ν−1

> 0,

where the last inequality is because the second term is greater than 1.

Thus, ζφX ,τ = 1 + ζφ∗D,τ > 1.

ζNX ,τ < −κ:

Given the mass of Home firms at Foreign NX = (φmin/φX)κNP where NP is the pool of Home

38

firms, we have

ζNX ,τ = −κζφX ,τ + ζNp,τ .

Since

NP =κ− (ν − 1)

κ

1

νWφκmin

[(ττ ∗)κF

κ−(ν−1)ν−1

φκDI

fD− φκXεI

∗

fX

(ττ ∗)κFκ−(ν−1)ν−1 − 1

],

we have

ζNp,τ =1

(ττ ∗)κFκ−(ν−1)ν−1

φκDI

fD− φκXεI

∗

fX

[(ττ ∗)κF

κ−(ν−1)ν−1

φκDI

fDκ(1 + ζφD,τ )−

φκXεI∗

fXκζφX ,τ

]

− κ(ττ ∗)κFκ−(ν−1)ν−1

(ττ ∗)κFκ−(ν−1)ν−1 − 1

.

Since

ζφD,τ = ζφX ,τ −τκF

κ−(ν−1)ν−1

τκFκ−(ν−1)ν−1 − (ρε)

κνν−1 (

f∗XfD

)κ−(ν−1)ν−1

,

we have

ζφD,τ + 1 = ζφX ,τ −τκF

κ−(ν−1)ν−1

τκFκ−(ν−1)ν−1 − (ρε)

κνν−1 (

f∗XfD

)κ−(ν−1)ν−1

+ 1

< ζφX ,τ .

Thus,

(ττ ∗)κFκ−(ν−1)ν−1

φκDI

fDκ(1 + ζφD,τ )−

φκXεI∗

fXκζφX ,τ <

[(ττ ∗)κF

κ−(ν−1)ν−1

φκDI

fD− φκXεI

∗

fX

]κζφX ,τ

and

ζNp,τ < κζφX ,τ −κ(ττ ∗)κF

κ−(ν−1)ν−1

(ττ ∗)κFκ−(ν−1)ν−1 − 1

< κζφX ,τ − κ.

Thus,

ζNX ,τ = −κζφX ,τ + ζNp,τ

< −κ.

39

(b) ζφX ,ε < −(1 + µ):

We obtain the elasticity of the exporting cutoff with respect to nominal exchange rates ε by

taking log (A.2) and differentiating it with respect to ε:

ζφX,ε = ζφ∗D,ε −ν

ν − 1,

where ζφ∗D,ε is the elasticity of the domestic cutoff level (φ∗D) with respect to ε for Foreign

firms.

We have ν/(ν − 1) = 1 + µ, and it can be shown that ζφ∗D,ε < 0.

Thus, ζφX ,ε = ζφ∗D,ε −νν−1

< −(1 + µ).

ζNX ,ε > κ(1 + µ):

We have

ζNX ,ε = −κζφX ,ε + ζNp,ε.

Since ζφX ,ε < −(1 + µ), we have −κζφX ,ε > κ(1 + µ).

Since

ζNp,ε =1

(ττ ∗)κFκ−(ν−1)ν−1

φκDI

fD− φκXεI

∗

fX

[(ττ ∗)κF

κ−(ν−1)ν−1

φκDI

fDκζφD,ε −

φκXεI∗

fX(κζφX ,ε + 1)

],

ζφD,ε > 0 and ζφX ,ε < −(1 + µ), we have ζNp,ε > 0.

Thus, ζNX ,ε > κ(1 + µ). �

Proposition (a) argues that a decrease in tariffs, charged by partners, causes a decrease

in the exporting cutoff and an increase in the mass of exporters (the EM of exports). In

particular, the mass of exporters increases by more than κ percent after a 1 percent reduction

in tariffs.

Proposition (b) argues that a decrease in nominal exchange rates—appreciation—causes

an increase in the exporting cutoff and a decrease in the mass of exporters. In particular,

the mass of exporters decreases by more than κ(1 + µ) percent after a 1 percent decrease in

nominal exchange rates (a 1 percent Home-currency appreciation). With the appreciation of

the Home currency and fixed wages, the relative cost of Home to Foreign labor—W/εW ∗—

increases and thus Home firms become less competitive in the Foreign market.

Furthermore, as implied by Chaney’s (2008) model and our model, the elasticity of the

40

mass of exporters with respect to tariffs/nominal exchange rates can be affected by the

elasticity of substitution between varieties ν and thus the markup µ = 1/(ν − 1). The

intuition is that, for the same percent change in tariffs or nominal exchange rates, if the

elasticity of substitution (ν) is lower, that is, if the markup (µ) is higher, then a change in

profits is larger. It leads to a larger change in the exporting cutoff and thus a larger change

in the mass of exporters. Thus, in the more differentiated sector, the elasticity of the mass

of exporters with respect to tariffs/nominal exchange rates can be larger.

A.2 Nominal and Real Exchange Rates

Figure A1 demonstrates the relationship between nominal and real exchange rates for China’s

selected trade partners from 2000/01 to 2006/12. Notably, due to data constraints, to cal-

culate the real exchange rates for the euro area we use producer prices for the euro area’s

CPI.

41

Figure A1: Nominal and Real Exchange Rates for Selected Trade Partners

7.8

88.

28.

48.

68.

8Yu

an/U

SD

2000 2001 2002 2003 2004 2005 2006Year

NER RER

United States

78

910

1112

Yuan

/Eur

o

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Euro Area

.06

.07

.08

.09

.1Yu

an/Y

en

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Japan

.006

.006

5.0

07.0

075

.008

.008

5Yu

an/W

on

2000 2001 2002 2003 2004 2005 2006Year

NER RER

South Korea

1213

1415

16Yu

an/P

ound

2000 2001 2002 2003 2004 2005 2006Year

NER RER

United Kingdoms

.24

.26

.28

.3.3

2Yu

an/T

WD

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Taiwan

56

78

Yuan

/CAD

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Canada

44.

55

5.5

66.

5Yu

an/A

UD

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Australia

2.1

2.15

2.2

2.25

2.3

Yuan

/MYR

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Malaysia

.12

.14

.16

.18

.2Yu

an/IN

R

2000 2001 2002 2003 2004 2005 2006Year

NER RER

India

.000

4.0

006

.000

8.0

01.0

012

Yuan

/IDR

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Indonesia

.17

.18

.19

.2.2

1.2

2Yu

an/T

HB

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Thailand

.65

.7.7

5.8

.85

.9Yu

an/M

XN

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Mexico

12

34

5Yu

an/B

RL

2000 2001 2002 2003 2004 2005 2006Year

NER RER

Brazil

Note: A decrease (increase) in nominal/real exchange rates means China’s appreciation (depreciation).

Source: IMF.

42

Do Exporters Respond to Both Tari s and Nominal Exchange ...€¦ · gin of exports can respond to...

Documents

Transcript of Do Exporters Respond to Both Tari s and Nominal Exchange ...€¦ · gin of exports can respond to...