NORC at the University of Chicago
The University of Chicago

Import Competition and the Great US Employment Sag of the 2000s
Author(s): Daron Acemoglu, David Autor, David Dorn, Gordon H. Hanson, and Brendan Price
Source: Journal of Labor Economics, Vol. 34, No. S1 (Part 2, January 2016), pp. S141-S198
Published by: The University of Chicago Press on behalf of the Society of Labor Economists andthe NORC at the University of Chicago
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Import Competition and the Great
US Employment Sag of the 2000s
Daron Acemoglu, Massachusetts Institute of Technology
and National Bureau of Economic Research

David Autor, Massachusetts Institute of Technology
and National Bureau of Economic Research

David Dorn, University of Zurich and Centre for Economic
Policy Research

Gordon H. Hanson, University of California, San Diego,
and National Bureau of Economic Research

Brendan Price, Massachusetts Institute of Technology
Even before the Great Recession, US employment growth was
unimpressive. Between 2000 and 2007, the economy gave back theconsiderable employment gains achieved during the 1990s, with ahistoric contraction in manufacturing employment being a prime
contributor to the slump. We estimate that import competition
from China, which surged after 2000, was a major force behind bothrecent reductions in US manufacturing employment and—throughinput-output linkages and other general equilibrium channels—
weak overall US job growth. Our central estimates suggest job
losses from rising Chinese import competition over 1999–2011 in
the range of 2.0–2.4 million.
We thank David Card, Alexandre Mas, Alireza Tahbaz-Salehi, and numerous
participants at the National Bureau of Economic Research conference titled“The Labor Market in the Aftermath of the Great Recession” for questions and[ Journal of Labor Economics, 2016, vol. 34, no. 1, pt. 2]© 2015 by The University of Chicago. All rights reserved. 0734-306X/2016/34S1-0011$10.00
Submitted September 17, 2013; Accepted May 14, 2015; Electronically published December 16, 2015

S141

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I. Introduction
During the last decade of the twentieth century—christened the “Roaring
Nineties” by Krueger and Solow ð2002Þ—the US labor market exhibited avigor not seen since the 1960s. Between 1991 and 2000, the employmentto-population ratio rose by 1.5 percentage points among men and by morethan 3 percentage points among women. Following 5 years of rapid wage
growth accompanied by minimal inflation, the national unemployment
rate in the year 2000 reached a nadir of 4.0%, its lowest level since 1969.
Just 1 year later, the US labor market commenced what Moffitt ð2012Þ
terms a “historic turnaround” in which the gains of the prior decade wereundone. Between 2001 and 2007, male employment rates lost all of their
ground attained between 1991 and 2000. The rapid increase of female
employment rates halted simultaneously.1 The growth rate of employment averaged only 0.9% between 2000 and 2007—that is, during the
7 years before the onset of the Great Recession—versus 2.6% between
1991 and 2000 ðfig. 1Þ.2
This pre–Great Recession US employment “sag” of the 2000s is widely
recognized but poorly understood.3 It coincides with a significant increasein import competition from China. Between 1990 and 2011, the share ofworld manufacturing exports originating in China increased from 2% to16% ðHanson 2012Þ. China’s export surge is the outcome of deep economic reforms in the 1980s and 1990s, which were reinforced by thecountry’s accession to the World Trade Organization ðWTOÞ in 2001

suggestions that improved the article. We are grateful to Christina Pattersonfor excellent research assistance. Dorn acknowledges funding from the Spanish
Ministry of Science and Innovation ðECO2010-16726 and JCI2011-09709Þ. Autorand Hanson acknowledge funding from the National Science Foundation ðgrantSES-1227334Þ. Price acknowledges financial support from the Hewlett Foundation. Autor, Acemoglu, and Price acknowledge support from the Alfred P. Sloan
Foundation ðgrant 2011-10-12Þ. Contact the corresponding author, David Autor,
at dautor@mit.edu. Information concerning access to the data used in this article isavailable as supplementary material online.
1
See http://www.bls.gov/ilc/#laborforce for data on the size and the employment rate of the working-age population.
2
The employment series plotted in fig. 1, as well as the employment statistics
provided later in this section, are derived from County Business Patterns ðCBPÞ.
As detailed below, CBP covers all US employment except for self-employedindividuals, employees of private households, railroad employees, agricultural production employees, and most government employees.
3
Moffitt ð2012Þ studies potential causes for the sag, including wage levels, age
structure, family structure, taxes, transfers, minimum wage policies, and population health. Only declining male wage rates are found to have substantial explanatory power. Yet this explanation leaves unanswered the question of why male wages
fell. The concurrence of falling wages and falling employment-to-population ratios
suggests an inward shift in labor demand.

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FIG. 1.—Changes in US manufacturing and nonmanufacturing employment,
1991–2011. Employment is computed in the CBP. Employment counts arenormalized to unity in 1991. A color version of this figure is available online.

ðNaughton 2007Þ. The country’s share in US manufacturing imports has
shown an equally meteoric rise from 4.5% in 1991 to 10.9% in 2001, before
surging to 23.1% in 2011. Simultaneously, after staying relatively constantduring the 1990s, US manufacturing employment declined by 18.7%
between 2000 and 2007 ðfig. 1Þ.4
In this article, we explore how much of the US employment sag of the
2000s can be attributed to rising import competition from China. Our
methodology builds on recent work by Autor, Dorn, and Hanson ð2013,2015Þ, as well as related papers by Autor et al. ð2014Þ, Bloom, Draca, and
Van Reenen ð2015Þ, and Pierce and Schott ð2015Þ. Akin to Pierce andSchott ð2015Þ, we begin our analysis with industry-level empirical specifications.5 This approach enables us to estimate the direct effect of exposure to Chinese import competition on industry employment at the US
4
Using CBP data, we calculate that US manufacturing employment was 17.0
million in 1991, 17.1 million in 2000, 13.9 million in 2007, and 11.4 million in 2011.
5
The North American Free Trade Agreement ðNAFTAÞ also contributed tochanges in US trade over our sample period. See McLaren and Hakobyan ð2012Þ
on NAFTA’s impacts on US employment patterns. More broadly, Ebenstein et al.
ð2014Þ examine the impact of trade in the form of offshoring on the wages of US
workers, finding that workers switching out of manufacturing experience relatively large wage declines.

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national level. Our direct industry-level employment estimates come fromcomparing changes in employment across four-digit manufacturing industries from 1991 to 2011 as a function of industry exposure to Chinese
import competition. The first part of our article shows that there is a sizable
and robust negative effect of growing Chinese imports on US manufacturing employment.
Quantitatively, our direct estimates imply that had import penetrationfrom China not grown after 1999, there would have been 560,000 fewer
manufacturing jobs lost through the year 2011. Actual US manufacturing
employment declined from 17.2 million workers in 1999 to 11.4 million in
2011, making the counterfactual job loss from the direct effect of greater
Chinese import penetration amount to approximately 10% of the realized
job decline in manufacturing.
These direct effects do not, however, correspond to the full generalequilibrium impact of growing Chinese imports on US employment, which
also encompasses several indirect channels through which rising exposure
to import competition may affect employment levels. One source of indirect effects, also studied by Pierce and Schott ð2015Þ, is industry inputoutput linkages. These linkages can create both positive and negative changes
in US industry labor demand, generating a net employment change that is
ambiguous in sign. If an industry contracts because of Chinese competition, it may reduce both its demand for intermediate inputs produced in
the United States and its supply of inputs to other domestic industries. An
industry may thus be negatively affected by trade shocks either to its domestic suppliers or to its domestic buyers. The sign of the “downstream
effect”—through which import exposure propagates downstream from a
supplying industry to its customers—is theoretically ambiguous: while tradecompetition may reduce the domestic supply of certain inputs, such reductions may be offset by the increased supply of imported inputs.6 By contrast, the “upstream effect”—whereby import exposure within an industry
propagates upstream to its suppliers—should have unambiguously contractionary consequences for the upstream industry.7
We use the US input-output table for 1992 to estimate the effects of
upstream and downstream import exposure for both manufacturing andnonmanufacturing industries. Our initial measure of the upstream ðrespec6
Trade shocks to an industry’s suppliers will have negative effects on thatindustry if, because of specific investments, existing supply relationships are moreproductive or are able to provide highly customized inputs as generally presumed
in the industrial organization literature on vertical integration ðe.g., Williamson
1975; Hart and Moore 1990Þ.
7
An earlier version of this article ðAcemoglu et al. 2014aÞ reversed the terminology of upstream and downstream effects. We have adopted the present terminology for consistency with common usage in the literature on input-output effects.

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tively, downstreamÞ effect for an industry, which sums over the direct import exposure experienced by all other industries using as weights theirshare in the total output demands of ðrespectively, their input supplies toÞ
the industry in question, captures this notion.8 Estimates from this exercise indicate sizable negative upstream effects while, consistent with theanticipated ambiguity of downstream effects, the downstream magnitudes
are imprecisely estimated and unstable in sign. Our preferred measure ofindirect trade shocks further accounts not only for shocks to an industry’s
immediate buyers or suppliers but also for the full set of input-output
relationships among all connected industries ðe.g., shocks to an industry’s
buyers, its buyers’ buyers, etc.Þ. Applying this direct plus full input-output
measure of exposure increases our estimates of trade-induced job losses for
1999–2011 to 985,000 workers in manufacturing alone and to 1.98 million
workers in the entire economy. Thus, interindustry linkages magnify theemployment effects of trade shocks, doubling the size of the impact withinmanufacturing and producing an equally large employment effect outsideof manufacturing.
Our second empirical strategy, which focuses on local labor markets, ismotivated by the fact that analysis at the level of national industries fails to
capture two other potentially important and opposing general equilibrium
channels. One such additional channel is a reallocation effect from growing trade with China, which works through the movement of factors of
production from declining sectors to new opportunities and potentially counteracts any negative direct or industry linkage effects. In both HeckscherOhlin and Ricardo-Viner models of international trade, stronger importcompetition for one sector reduces the relative price of its final good andinduces the reallocation of labor and capital to sectors whose relative prices
have increased ðFeenstra 2003Þ. Under fully inelastic labor supply, no labormarket frictions, and other neoclassical assumptions that ensure that theaggregate economy is always at full employment, reallocation effects would,
by definition, exactly offset direct, upstream, and downstream effects so as torestore full employment. However, with imperfections in labor and othermarkets, there is no guarantee that reallocation effects will be sufficient torestore employment to the same level that would have emerged in the absenceof trade growth from China.
An additional general equilibrium channel operates through aggregate
demand effects, multiplying the negative direct and indirect effects ofimport growth from China. Through familiar Keynesian-type multipliers,
domestic consumption and investment may be depressed, extending emSee Long and Plosser ð1983Þ and Acemoglu et al. ð2012Þ for the reasoning
behind this value share definition, which also corresponds to the relevant entries in
the input-output tables. A detailed derivation is provided in app. B.
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ployment losses to sectors not otherwise exposed to import competition. A
negative effect of increased import competition on aggregate demandnecessarily requires that employment reallocation in response to a negativetrade shock is incomplete, such that aggregate earnings decline, and thisdecline is multiplied throughout the economy via demand linkages.
We jointly estimate reallocation and aggregate demand effects ðin netÞ atthe level of local labor markets by exploiting the impact of trade shocks
within US commuting zones ðCZsÞ. If the reallocation mechanism isoperative, then when an industry contracts in a CZ as a result of Chinesecompetition, some other industry in the same labor market should expand.
Some component of aggregate demand effects should also take place within
local labor markets, as shown by Mian and Sufi ð2014Þ in the context of therecent US housing bust: if increased trade exposure lowers aggregate employment in a location, reduced earnings will decrease spending on nontraded
local goods and services, magnifying the impact throughout the local economy. Because aggregate demand effects also have a national component,
which our approach does not capture, focusing on local labor markets islikely to provide a lower bound on the sum of reallocation and aggregate
demand effects.9
Empirically, our second strategy examines changes in employment in
CZs that have different levels of exposure to Chinese competition by virtueof differences in their initial pattern of industrial specialization, a strategy
also used by Autor et al. ð2013Þ. The reallocation effect should result in agreater expansion of employment in nonexposed industries—meaning
nontradable industries as well as tradable industries not significantly exposed to trade with China. Surprisingly, we find no robust evidence for thiseffect: the estimated impact of import competition on employment in nonexposed industries is very modest in magnitude and statistically indistinguishable from zero. The reallocation of employment into nonexposed industries appears to be swamped by the adverse effect of the aggregate
demand channel, which presumably inhibits labor reabsorption.
Our estimates of local general equilibrium effects imply that import
growth from China between 1999 and 2011 led to an employment reduction of 2.4 million workers, inclusive of employment changes within
9

Of course, reallocation effects may also have a national component due to the
movement of labor across regions. As we discuss in Sec. II, in practice there
appears to be little response of local labor supply to location-specific increases inimport competition from China ðAutor et al. 2013, 2014Þ, leading us to view
reallocation effects as being primarily local in nature. Another complicating factor
is that, in the presence of labor and product market imperfections, the decline of anindustry in the local labor market may lead to the expansion of some tradable
industries in other labor markets, making the local reallocation effects a lower
bound on the aggregate reallocation effects.

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nonexposed sectors. Consistent with the idea that import competition
may have negative general equilibrium effects on local employment, this figure exceeds our national industry-level estimate of the direct and indirect
disemployment effects of rising import exposure mentioned above. As
noted below, neither the CZ-level nor the national estimate fully incorporates all of the adjustment channels encompassed by the other. The national industry estimates exclude reallocation and aggregate demand effects,
whereas the CZ estimates exclude the national component of these two
effects, as well as the nonlocal component of input-output linkage effects.
Because the CZ-level estimates suggest that general equilibrium forces
magnify rather than offset the effects of import competition, we view ourindustry-level estimates of employment reduction as providing a conservative lower bound.
Our analysis of the aggregate employment consequences of importcompetition builds on the recent work of Autor et al. ð2013, 2015Þ by expanding their CZ-level analysis to include analysis at the level of national
industries, a dimension they do not consider, and by characterizing the alternative mechanisms—reallocation versus changes in aggregate demand—
through which trade induces employment decline at the local level. Our
national industry approach is similar in spirit to that of Bloom et al. ð2015Þ
and Pierce and Schott ð2015Þ. Pierce and Schott, in particular, explore how
China’s 2001 World Trade Organization accession affected US manufacturing employment. Our article, while complementary to theirs, differs in two
respects. The first is in terms of identification strategy. Whereas Pierce andSchott seek to identify the growth in China trade that resulted from the post2001 removal of uncertainty surrounding China’s most-favored-nation accessto the US market, our identification strategy captures China’s trade growth
due to broader productivity-driven changes in its export supply. Further,
our article expands the analysis to include the transmission of trade shocks
to nonmanufacturing sectors and the estimation of employment effects resulting from reallocation across sectors and changes in aggregate demand.
We begin in Section II by outlining the conceptual framework that motivates our empirical analysis. Section III describes our empirical approachto estimating the effects of exposure to trade shocks and briefly discusses
the data. Section IV gives our primary ordinary least squares ðOLSÞ andtwo-stage least squares ð2SLSÞ estimates of the impact of trade shocks on
employment and also considers additional labor market outcomes. Section V expands the analysis to include intersectoral linkages. Section VI
presents estimation results for data on local labor markets. Section VII concludes the article. Appendix A contains additional empirical results and robustness checks, and appendix B contains the derivation of our upstream
and downstream import exposure measures from a simple general equilibrium model with input-output linkages.

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II. Conceptual Framework
We start with a brief outline of the conceptual framework that motivates
our empirical work. Consider a simple decomposition of the total national
employment impact of increased Chinese trade exposure:10
National employment impact
5 Direct impact on exposed industries
1 Indirect impact on linked industries
1 Aggregate reallocation effects
1 Aggregate demand effects:
Here, the direct impact is the reduction in employment in industries whose
outputs compete with imports from China. Added to this direct effect is
an indirect effect arising because other industries linked to the affectedindustry through the input-output matrix are also likely to see changes in
output.11 For example, the chemical and fertilizer mining industry—whichis in nonmanufacturing—sells 74% of its output to the manufacturingsector. Its largest single manufacturing customer is industrial organic
chemicals not elsewhere classified, which accounts for 15% of its sales.
Similarly, the iron and ferroalloy ores industry sells 83% of its output tothe manufacturing sector, two-thirds of which goes to the blast furnace andsteel mill industry. Accordingly, a shock to the demand for a given domestic
manufactured good is likely to indirectly affect demand for, and reduce employment in, industries that supply inputs to the affected industry, whether
in manufacturing or nonmanufacturing. We refer to these linkages as upstream effects, by which industries exposed to import competition indirectly affect industries that are located upstream of them in input-output
space.12
Conversely, a trade shock to the suppliers of a given industry ðe.g., thesuppliers of tires to the automobile industryÞ may also affect the industries
that are its customers. The direction of this effect is generally ambiguous.
On the one hand, from the perspective of purchasing industries, the trade
shock expands input supply and puts downward pressure on input prices
and thus may tend to expand employment in the industries that consume

10
We follow the standard practice in such decompositions and fold the “covariance” terms into the “main effects” ðso that the magnitudes are not independentof the order in which these different terms are evaluatedÞ.
11
See, among others, Long and Plosser ð1983Þ and Acemoglu et al. ð2012Þ on the
propagation of shocks through the input-output network of the economy.
12
Unfortunately, the terminology of upstream and downstream effects is open
to confusion, since upstream effects—i.e., effects that propagate upstream—workthrough the import exposure experienced by downstream industries, and similarly
for downstream effects.

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these inputs ðGoldberg et al. 2010Þ.13 On the other hand, the trade shock
may destroy existing long-term relationships for specialized inputs as domestic input suppliers are driven out of business, creating a force toward
contraction in the industries that were their customers. We refer to suchlinkages as downstream effects, since they propagate from an importexposed industry to industries located downstream in the production chain.
We estimate these effects on linked industries using the input-output matrixof the US economy as described below.
We begin our empirical analysis with industry-level regressions that estimate the direct impact of import competition on employment in exposedindustries ðSec. IVÞ and subsequently add the indirect employment impacts arising from input-output linkages between industries ðSec. VÞ. The
industry-level analysis thus captures the first two components of the aggregate national employment effect, the direct impact on exposed industries plus the indirect impact on linked industries. The industry-level regressions do not, however, encompass the third and the fourth componentsof the national employment effect: the reallocation effect, which captures
the potential increase in employment from the expansion of other industries to absorb the factors of production freed by contracting industries, and the aggregate demand effect, which corresponds to the impactof Keynesian-type multipliers operating through local or national shifts in
consumption and investment.14
To obtain estimates of the magnitudes of these two additional effects,
we turn in Section VI to local labor market analysis, focusing on the employment impact of increased import competition from China at the CZ
level. The total employment effect observed in a local labor market can be
decomposed as
Local employment impact
5 Direct impact on exposed industries
1 Local impact on linked industries
1 Local real location effects 1 Local demand effects:
13
Consistent with this reasoning, De Loecker et al. ð2014Þ find substantial
negative domestic product price effects from trade liberalization in India, andGoldberg et al. ð2010Þ document that greater availability of imported intermediate
inputs is associated with more rapid introduction of new product varieties bydomestic firms, also in the Indian context.
14
It is in theory possible for the aggregate demand effect to be positive; forinstance, aggregate demand may increase because the aggregate price level declines
as a result of the lower costs of imported products from China. We view this positive
channel as second-order and in general presume that the aggregate demand effect,
working in the standard Keynesian fashion, amplifies the potential negative direct
impact of trade shocks. This is consistent with the results from our local labormarket analysis, which indicate that the sum of reallocation and demand effects isnegative.

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We hypothesize that the direct impact at the local level, when scaled appropriately by the size of the industry in the local labor market, is comparable to the direct impact estimated at the national level. The other threeeffects could potentially differ between the local and the aggregate levels.
For instance, even though linked industries tend to co-locate ðe.g., Ellison,
Glaeser, and Kerr 2010Þ, only part of the input-output linkages will bewithin the same local labor market, and the local impact on linked industries may thus be much smaller than the aggregate effect.
What makes our local labor market analysis informative is that local
reallocation and local demand effects are linked to their aggregate counterparts. Consider the reallocation effects first. Local labor markets are aplausible unit of analysis for the study of this channel. As a local labormarket experiences a loss of jobs when local industries contract in response
to rising import competition, there should be an adjustment of quantities
within the same labor market, despite the fact that prices are, at least in part,
determined in the national or the international equilibrium. If the extent ofworker migration between local labor markets in response to these labormarket shocks is modest, as suggested by the evidence in Autor et al. ð2013,2014Þ and Notowidigdo ð2013Þ, this adjustment will take the form of reallocation from declining industries to others within this locale.15
An important component of aggregate demand effects also plausibly
takes place within local labor markets. Mian and Sufi ð2014Þ show thatduring the Great Recession, US counties suffering large wealth losses
because of particularly severe declines in housing values also saw large
declines in employment, consistent with local transmission of shocks toaggregate demand. Components of the aggregate demand effect that operate at the national level will not be captured by our analysis, however, as they
will be common across locations. Our empirical strategy seeks to identify
the combined impact of reallocation and aggregate demand effects by quantifying how trade-induced shocks have an impact on a CZ’s employment
in nonexposed industries—defined as industries that are not exposed toimports from China either through direct product market competition or
through interindustry purchases of intermediate inputs.
Overall, this discussion suggests that our local labor market strategy
will provide an informative alternative estimate of the aggregate employment impact of greater import competition from China, though this is
likely to be an underestimate of the aggregate effects because it ignores partof the impact on linked industries and also excludes demand effects thathave no counterpart at the local level. In what follows, we will separately
Complementing this US-based evidence, Balsvik, Jensen, and Salvanes ð2014Þ
and Dix-Carneiro and Kovak ð2015Þ document weak labor mobility responses totrade-induced employment shocks in Norway and Brazil, respectively. As discussed in footnote 9, there are some components of reallocation that might takeplace outside the local labor market.
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compute the implied aggregate effects consisting of the sum of the direct
impact and the impact on linked industries from our national industrylevel analysis, and the total employment impact from the local analysis.
III. Empirical Approach
Sweeping economic reforms initiated in the 1980s and extended in the
1990s permitted China to experience rapid industrial productivity growthðNaughton 2007; Hsieh and Ossa 2011; Zhu 2012Þ, rural to urban migration
flows in excess of 150 million workers ðLi et al. 2012Þ, and massive capital
accumulation ðBrandt, Van Biesebroeck, and Zhang 2012Þ, which together
caused manufacturing to expand at a breathtaking pace. What did thisgrowth mean for US employment inside and outside manufacturing? We
seek to capture the changes in US industry employment induced by shifts
in China’s competitive position and the subsequent increase in its exports,
accounting for input-output linkages between industries and other indirect channels of transmission. We subsequently consider how these labordemand shifts can be aggregated to national totals.
A. Industry Trade Shocks
Our baseline measure of trade exposure is the change in the importpenetration ratio for a US manufacturing industry over the period 1991–
2011, defined as
DIPjt 5D MUC
j;t
Yj;91 1 Mj;91 2 Ej;91;

ð1Þ

where for US industry j, DMUC
j;t is the change in imports from China overthe period 1991–2011 ðwhich in most of our analysis we divide into two
subperiods, 1991–99 and 1999–2011Þ and Yj,91 1 Mj,91 2 Ej,91 is initialabsorption ðmeasured as industry shipments, Yj,91, plus industry imports,
Mj,91, minus industry exports, Ej,91Þ. We choose 1991 as the initial year as itis the earliest period for which we have the requisite disaggregated bilateral trade data for a large number of country pairs that we can match to
US manufacturing industries.16 The quantity in ð1Þ can be motivated bytracing export supply shocks in China—due, for example, to productivity
growth—through to demand for US output in the markets in which theUnited States and China compete. Supply-driven changes in China’s exports will tend to reduce demand for and employment in US industries.
16

Our empirical approach requires data not just on US trade with China but
also on China’s trade with other partners. Specifically, we require trade data reported under Harmonized System ðHSÞ product codes in order to match with US
Standard Industrial Classification ðSICÞ industries. The year 1991 is the earliest inwhich many countries began using the HS classification.

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One concern about ð1Þ as a measure of trade exposure is that observedchanges in the import penetration ratio may in part reflect domestic shocks
to US industries that affect US import demand. Even if the dominant factors
driving China’s export growth are internal supply shocks, US industry
import demand shocks may still contaminate bilateral trade flows. To
capture this supply-driven component in US imports from China, we instrument for trade exposure in ð1Þ with the variable
DIPOjt 5

DMOCj;tYj;88 1 Mj;88 2 Xj;88;

ð2Þ

where DMOC
j;t is the growth in imports from China in industry j during
the period t ðin this case 1991–2011 or some subperiod thereofÞ in eight
other high-income countries excluding the United States.17 The denominator in ð2Þ is initial absorption in the industry in 1988. The motivation for
the instrument in ð2Þ is that high-income economies are similarly exposed
to growth in imports from China that is driven by supply shocks in the
country. The identifying assumption is that industry import demand
shocks are uncorrelated across high-income economies and that there are
no strong increasing returns to scale in Chinese manufacturing ðwhichmight imply that US demand shocks will increase efficiency in the affected
Chinese industries and induce them to export more to other high-income
countriesÞ.18
Figure A1 ðin app. AÞ plots the value in ð1Þ against the value in ð2Þ forall US manufacturing industries at the four-digit level, as defined below,
which is equivalent to the first-stage regression in our subsequent estimation without detailed controls. The coefficient is 0.98 and the t-statistic
and R-squared are 7.0 and .62, respectively, indicating the strong predictive
power of import growth in other high-income countries for US import
growth from China.19
17

These countries are Australia, Denmark, Finland, Germany, Japan, New
Zealand, Spain, and Switzerland, which represent all high-income countries forwhich we can obtain disaggregated bilateral trade data at the HS level back to 1991.
18
See Autor et al. ð2013, 2014Þ for further discussion of threats to identification
using this instrumentation approach.
19
Modeling the China trade shock as in eq. ð1Þ does not exclude the role of global
production chains. During the 1990s and 2000s, approximately half of China’s manufacturing exports were produced by export processing plants, which import partsand components from abroad and assemble these inputs into final export goodsðFeenstra and Hanson 2005Þ. Our instrumental variable strategy does not require
China to be the sole producer of the goods it ships abroad; rather, we require thatthe growth of its gross manufacturing exports is driven largely by factors internal to
China ðas opposed to shocks originating in the United StatesÞ, as would be the caseif, plausibly, the recent expansion of global production chains involving China is
primarily the result of its hugely expanded manufacturing capacity.

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A potential concern about our analysis is that we largely ignore US exports to China, focusing primarily on trade flows in the opposite direction.
This is for the simple reason that our instrument, by construction, has little predictive power for US exports to China. Nevertheless, to the extent
that our instrument is valid, our estimates will correctly identify the direct
and indirect effects of increased import competition from China in particular because there is no reason for trade to balance at the industry or
region level, so we do not need to simultaneously treat exports to China in
our analysisÞ. We also take comfort from the fact that imports from China
are much larger—approximately five times as large—than manufacturingexports from the United States to China ðfig. 2Þ.20
B. Data Sources
Data on international trade for 1991–2011 are from the UN Comtrade
Database ðhttp://comtrade.un.org/db/default.aspxÞ, which gives bilateralimports for six-digit Harmonized Commodity Description and Coding
System ðHSÞ products. To concord these data to four-digit Standard Industrial Classification ðSICÞ industries, we first apply the crosswalk in
Pierce and Schott ð2012Þ, which assigns 10-digit HS products to four-digit
SIC industries ðat which level each HS product maps into a single SIC
industryÞ, and aggregate up to the level of six-digit HS products and fourdigit SIC industries ðat which level some HS products map into multiple
SIC industriesÞ. To perform this aggregation, we use data on US importvalues at the 10-digit HS level, averaged over 1995–2005. The crosswalk
assigns HS codes to all but a small number of SIC industries. We therefore
slightly aggregate the four-digit SIC industries so that each of the resulting
397 manufacturing industries matches to at least one trade code and none
is immune to trade competition by construction. To ensure compatibility
with the additional data sources below, we also aggregate together a few
additional industries such that our final data set contains 392 manufacturing industries. All import amounts are inflated to 2007 US dollars using
the Personal Consumption Expenditure ðPCEÞ deflator.
Our main source of data on US employment is County Business Patterns ðCBPÞ for the years 1991, 1999, 2007, and 2011. CBP is an annual
data series that provides information on employment, firm size distribu20

A second rationale for our import focus is data constraints. Many US exports
to China are in the form of indirect exports via third countries or embodied services
of intellectual property, management expertise, or other activities involving skilledlabor. These indirect and service exports are difficult to measure because the directexporter may be a foreign affiliate of a US multinational or because they occur via achain of transactions involving third countries. As such exports tend to be intensive
in highly skilled labor, they may have only modest direct impacts on the employment of production workers, though their indirect impacts are difficult to gauge
with available data.

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Acemoglu et al.

FIG. 2.—Bilateral US-China trade flows and Chinese import penetration, 1991–
2011. Trade data are taken from the UN Comtrade Database. Imports and exports
are deflated to 2007 US dollars using the PCE price index. Chinese import penetration is constructed by dividing US manufacturing imports from China by US
domestic manufacturing absorption, defined as US domestic manufacturing output plus imports less exports. Export data are available only from 1992 onward.
The import penetration ratio series ends in 2009 because computing the denominator requires use of the NBER-CES Manufacturing Industry Database, which
ends in 2009. A color version of this figure is available online.

tion, and payroll by county and industry. It covers all US employment
except self-employed individuals, employees of private households, railroad employees, agricultural production employees, and most government employees.21
To supplement the employment and establishment count measuresavailable from the CBP, we utilize the NBER—Center for Economic
Studies Manufacturing Industry Database for the years 1971–2009 ðthe21
CBP data are extracted from the Business Register, a file of all known US
companies that is maintained by the US Census Bureau; see http://www.census.gov/econ/cbp/index.html. To preserve confidentiality, CBP information on employment by industry is sometimes reported as an interval instead of an exact count.
We compute employment in these cells using the fixed-point imputation strategy
developed by Autor et al. ð2013Þ.

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Import Competition and the Great US Employment Sag

S155

latter being the latest year availableÞ.22 These data allow us to explore labormarket outcomes not reported in the CBP, as well as to perform a falsification exercise not possible in the CBP. We additionally draw on the
NBER-CES data to compute measures of the production structure in eachindustry, subsequently used as controls, including production workers as ashare of total employment, the log average wage, the ratio of capital to value
added, computer investment as a share of total investment, and high-techequipment as a share of total investment. Additionally, we create industry
pretrend controls for the years 1976–91, including the changes in industry
log average wages and in the industry share of total US employment.
A final data source used in our analysis is the 1992 input-output table
for the US economy ðfrom the US Bureau of Economic Analysis, http://
www.bea.gov/industry/io _ benchmark.htmÞ, which we use to trace upstream and downstream demand linkages between industries both insideand outside of US manufacturing. We discuss our application of inputoutput tables in more detail below.
IV. Estimates of the Direct Impact of Trade
Exposure on Employment
We begin by estimating the direct effect of trade exposure on employment
over the period 1991–2011 using aggregate, industry-level regressions.
A. Baseline Results for National Industries
Our initial specification has the following form:
DLjt 5 at 1 b1 DIPjt 1 gXj0 1 ejt ;ð3Þ

where DLjt is 100 times the annual log change in employment in industry jover time period t; DIPjt is 100 times the annual change in import penetration from China in industry j over period t as defined in ð1Þ; Xj0 is a setof industry-specific start-of-period controls ðspecified laterÞ; at is a periodspecific constant; and ejt is an error term. We fit this equation separately forstacked first differences covering the two subperiods 1991–99 and 1999–
2011, where in some specifications we shorten the second subperiod to1999–2007 in order to evaluate employment impacts prior to the onset of
the Great Recession. Variables specified in changes ðdenoted by DÞ are
annualized since equation ð3Þ is estimated on periods of varying lengths.
The elements in the vector of controls Xj0, when included, are each normalized with mean zero so that the constant term in ð3Þ reflects the change
22

The NBER-CES database contains annual industry-level data from 1958–
2009 on output, employment, payroll and other input costs, investment, capitalstocks, total factor productivity, and various industry-specific price indexesðBecker, Gray, and Marvakov 2013Þ. Data and documentation are at http://www.nber.org/data/nberces5809.html.

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Acemoglu et al.

in the outcome variable conditional only on the variable of interest, DIPjt.
Most outcome variables are measured at the level of 392 four-digit manufacturing industries, while later models also estimate spillovers to 87 nonmanufacturing industries. Regression estimates are weighted by startof-period industry employment, and standard errors are clustered at thethree-digit industry level to allow for arbitrary error correlations withinlarger industries over time.23
Table 1 summarizes the import exposure and employment variables
used in initial estimates of equation ð3Þ. The employment-weighted mean
industry saw Chinese import exposure rise by 0.5 percentage points per
year between 1991 and 2011, with more rapid penetration during 1999–
2007 than during 1991–99: 0.8 versus 0.3 percentage points, respectively.
Growth from 2007 to 2011, at 0.3 percentage points per year, indicates amarked slowdown in import expansion in the late 2000s. The slowdown
during that period is the combined effect of a steep decline in US trade in2008 and 2009 and an equally dramatic recovery in 2010 ðLevchenko, Lewis,
and Tesar 2010Þ, which together left import penetration rates modestly
higher.24
Changes in import penetration are highly right skewed across manufacturing industries, with the mean increase exceeding the median by afactor of 3.5. We find a similar pattern of import penetration change andskewness in the other high-income countries used to construct the importpenetration instrument, where this skewness reflects China’s strong comparative advantage in labor-intensive industries. Table 1 also shows that the
manufacturing decline accelerated throughout the sample: the averageindustry contracted by 0.3 log points per year between 1991 and 1999, by
3.6 log points per year between 1999 and 2007, and by 5.7 log points per
year in the final period 2007–11. The within-industry growth rate of nonmanufacturing employment also slowed across the three subperiods of oursample, but the deceleration was not nearly as pronounced as in manufacturing.
Table 2 presents a simple stacked first-difference model for the two time
periods 1991–99 and 1999–2011, with the change in import penetration
and a dummy for each time period as the only regressors. Alongside theseestimates, we also present results from stacking the time periods 1991–99
23

There are 135 three-digit manufacturing industry clusters encompassing the392 four-digit industries. Because our nonmanufacturing data have already been
extensively aggregated to 87 industries for concordance with the BEA input-output
table, we treat each of the 87 nonmanufacturing industries as a single cluster.
24
Explanations for the excess sensitivity of trade flows during the Great Recession include the role of shocks to the credit market and trade finance ðAmiti and
Weinstein 2011; Chor and Manova 2012Þ and to global production networksðLevchenko et al. 2010Þ. Other explanations dwell on the large drop in durable good
spending during the crisis ðEaton et al. 2013Þ.

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392

392
87

Instrument for D in US exposure to Chinese imports

100  annual log D in employment:
Manufacturing industries

Nonmanufacturing industries

22.71
ð3.07Þ
1.33
ð1.46Þ

.50
ð.94Þ
.44
ð.76Þ

Mean/SD

1.02

22.05

.15

.14

Median

4.62
5.75

25.73

8.59

10.93

Max

238.32

2.52

2.02

Min

2.30
ð3.49Þ
2.46
ð2.38Þ

.27
ð.75Þ
.18
ð.44Þ

1991–99
Mean/SD

24.32
ð3.85Þ
.57
ð1.56Þ

.66
ð1.33Þ
.60
ð1.07Þ

1999–2011
Mean/SD

23.62
ð4.15Þ
1.54
ð1.59Þ

.84
ð1.61Þ
.60
ð1.07Þ

1999–2007
Mean/SD

25.73
ð5.02Þ
21.37
ð2.83Þ

.30
ð1.68Þ
.62
ð1.32Þ

2007–11
Mean/SD

NOTE.—For each manufacturing industry, the change in US exposure to Chinese imports is computed by dividing 100  the annualized increase in the value of US imports over theindicated period by 1991 US market volume in that industry. The instrument is constructed by dividing 100  the annualized increase in imports from China in a set of comparison
countries by 1988 US market volume in the industry. The quantities used in these computations are deflated to constant dollars using the PCE price index. Employment changes arecomputed in the CBP. All observations are weighted by 1991 industry employment.

392

100  annual D in US exposure to Chinese importsN

1991–2011

Table 1
Industry-Level Changes in Chinese Import Exposure and US Manufacturing Employment

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OLS

2.30
ð.37Þ
24.32***
ð.37Þ

OLS

2.81***
ð.16Þ
2.08
ð.36Þ
23.79***
ð.33Þ

ð2Þ

2SLS

21.30***
ð.41Þ
.05
ð.36Þ
23.46***
ð.33Þ

ð3Þ

Stacked Differences ðN 5 784Þ2SLS

22.58***
ð.38Þ

21.24***
ð.37Þ
.04
ð.36Þ

1991–2007
ð4Þ

.32
ð.43Þ
2SLS

22.30**
ð1.12Þ

1991–99
ð5Þ

23.55***
ð.34Þ
2SLS

21.16***
ð.37Þ

1999–2011ð6Þ

22.68***
ð.39Þ
2SLS

21.12***
ð.34Þ

1999–2007
ð7Þ

Separately by Period ðN 5 392Þ

21.96***
ð.27Þ
2SLS

21.49***
ð.47Þ

1991–2011
ð8Þ

NOTE.—Columns 1– 4 report results from stacking log employment changes and changes in US exposure to Chinese imports over the periods 1991–99 and either 1999–2011
or 1999–2007, as indicated ðN 5 784 5 392 four-digit manufacturing industries  two periodsÞ. Columns 5–8 report results from regressing the employment change over theindicated period on the change in US exposure to Chinese imports over the same period ðN 5 392Þ. Employment changes are computed in the CBP and are expressed as 100 
annual log changes. In 2SLS specifications, the change in US import exposure is instrumented as described in the text. In all specifications, observations are weighted by 1991
employment. Standard errors in parentheses are clustered on 135 three-digit industries in all specifications.
** p < .05.
*** p < .01.

Estimation method

Constant

1f1999–2007g

1f1999–2011g

1f1991–99g

100  annual D in US exposure
to Chinese imports

1991–2011
ð1Þ

Table 2
Effect of Import Exposure on Log Employment in US Manufacturing Industries: OLS and 2SLS Estimates

Import Competition and the Great US Employment Sag

S159

and 1999–2007 and from fitting the model separately for the three subperiods 1991–99, 1999–2011, and 1999–2007. These additional specifications permit inspection of results before and after the commencement of
the 2000s US employment sag and allow for comparison of the results forthe 2000s with and without including the Great Recession years. We alsopresent results for the single long difference, 1991–2011, for comparison
against the stacked first differences.
In column 1, which excludes the import penetration variable, the time
dummies reflect the ðemployment-weightedÞ mean annual within-industry
change in employment in each period. Column 2 adds the observed import
exposure measure without instrumentation. This variable is negative andhighly significant, consistent with the hypothesis that rising import penetration lowers domestic industry employment. Nevertheless, as noted
above, this OLS point estimate could be biased because growth in importpenetration is driven partly by changes in domestic supply and demand.
Column 3 mitigates this simultaneity bias by instrumenting the observedchanges in industry import penetration with contemporaneous changes inother-country China imports as specified in equation ð2Þ above. The estimate in column 3 implies that a 1 percentage point rise in industry importpenetration reduces domestic industry employment by 1.3 percentage
points ðt-ratio of 3.2Þ. Column 4, which stacks the periods 1991–99 and
1999–2007, shows that the coefficient of import penetration is very similar
if we restrict attention to the years preceding the Great Recession.
The remaining columns of table 2 present bivariate estimates of thisrelationship separately by subperiod. The coefficient on trade exposure isnegative and statistically significant in all time periods and is largest in
absolute value for 1991–99 and smallest for 1999–2007. Even though thesensitivity of employment to import penetration is greater before 2000,
the much faster growth in China’s imports after 2000 produces an overall
impact of trade on employment that, as we discuss below, is considerablylarger in the latter period. The sensitivity of employment to trade for 1999–
2011 is similar to the estimate for 1999–2007, despite the onset of the global
financial crisis in 2007 and the associated dislocation of worldwide tradepatterns.25
A simple long-difference model for the change in manufacturing employment over the full 1991–2011 period ðcol. 8Þ also supports a negative
relationship between import penetration and US manufacturing employ-25

In the United States, imports plus exports divided by GDP fell by a stunning
22% from the first quarter of 2008 to the first quarter of 2009. However, imports
fully recovered in 2010 and continued to grow in 2011. The exaggerated cyclicalswings in trade surrounding the Great Recession thus mix with the continued
secular growth in China’s exports to the United States over the period.

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ment. The coefficient estimates in column 3, for the stacked first differences, and column 8, for the long time difference, are quite similar, reflecting strong persistence in the growth in China’s import penetration within
industries. Replacing stacked first differences with the long difference may
remove cyclical variation in the data, accounting for the mildly larger coefficient estimates in the latter case.
Returning to the results in column 3 of table 2, we evaluate the economic
magnitude of these estimates by constructing counterfactual changes in
employment that would have occurred in the absence of increases in
Chinese import competition. Using equation ð3Þ, we write the difference
between actual and counterfactual manufacturing employment in year t as∼

DLcft 5 o Ljt ð1 2 e2b1 DIPjt Þ;
^

ð4Þ

j

^1 is the 2SLS coefficient estimate from ð3Þ and DIP
fjt is the increase
where bin import penetration from China that we attribute to China’s improving
competitive position in industry j between 1991 ðor 1999Þ and year t.
fjt by multiplying the obFollowing Autor et al. ð2013Þ, we estimate DIP
served increase in import penetration DIPjt with the partial R-squared
from the first-stage regression of ð1Þ on the instrument in ð2Þ, which has
a value of 0.56 in our baseline specification in column 3 in table 2. When
our instrument is valid and there is no measurement error, this partial
fjt variable is a consistent estimate of the contriR-squared adjusted DIP
bution of Chinese import supply shocks to changes in import penetration.
In constructing the counterfactuals, we further assume that all other factors, including observed covariates and unobserved shocks captured by theerror term in ð3Þ, would be unaffected by the artificially imposed reductionin the growth of import penetration from China.
We collect these counterfactual estimates in table 8 below, where wecompare employment estimates across three different estimation strategies.
The first row of table 8 reports counterfactual employment differences
implied by the estimates in table 2, where we evaluate changes for 1991–99,
1999–2011, and the entire 1991–2011 period. Using coefficient estimatesfrom column 3, we calculate that had import penetration from China remained unchanged between 1991 and 2011, manufacturing employment
would have fallen by 837,000 fewer jobs over the full 1991–2011 spanand by 560,000 fewer jobs during the employment sag era of 1999–2011.
Observed manufacturing employment changes over these time periods
were 25.6 million workers ð11.4 million 2 17.0 millionÞ and 25.8 million
workers ð11.4 million 2 17.2 millionÞ, respectively. The larger quantity forthe second period is indicative of the modest growth in manufacturing
employment of 200,000 workers that occurred between 1991 and 1999. By
shutting down China’s import growth, the contraction of US manufac-

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Import Competition and the Great US Employment Sag

S161

turing employment suggested by our estimates would have been 14.9 percentage points less over 1991–2011 and 9.7 percentage points less for theperiod after 1999. It is also worth noting that counterfactual reductions
in employment for the period 1991–2007—based on the specification incolumn 4 of table 2—amount to 853,000, quite similar to our estimates for
1991–2011.
B. Comparison to Other Estimates in the Literature
How do our estimates of the direct effect of import competition on manufacturing employment compare with those found in the literature? There
are few estimates to consider, as the majority of work on the labor market
implications of globalization addresses not the absolute employment effects of trade but its impact on relative wages and relative employment
levels by skill ðe.g., Harrison, McLaren, and McMillan 2011Þ. Trade impacts on absolute employment levels are a less common object of study,
perhaps reflecting modeling conventions that impose inelastic labor supply
and full employment.
In an influential treatment of trade impacts on US manufacturing,
Bernard, Jensen, and Schott ð2006Þ estimate that import penetration from
low-income countries—with China being the largest member of thisgroup by far—accounts for 14% of the total decline in manufacturing
employment of 675,000 workers that occurred between 1977 and 1997.26
Their specification differs from ours, making a direct comparison of thetwo sets of results difficult to perform. They regress the change in log
employment at the level of the manufacturing plant ðrather than industryÞ
on the initial level ðrather than changeÞ of the share of low-income countries in industry imports ðrather than the import penetration rateÞ. Despite
these differences, Bernard et al. find a relatively high sensitivity of employment to import competition. But over their period of study, the annual
increase in import penetration from low-income countries in US manufacturing was only 0.09 percentage points,27 whereas over our sample period
the annual increase in import penetration from China alone was 0.50 per-26
In related work, Artuç, Chaudhuri, and McLaren ð2010Þ evaluate how costs
to workers of moving between sectors dampen the employment response tochanges in trade barriers, and Muendler and Becker ð2010Þ and Harrison and
McMillan ð2011Þ estimate the responsiveness of employment in multinational
companies to changes in foreign wages. This work tends to emphasize the elasticity of employment with respect to changes in trade barriers or foreign production costs, rather than producing estimates of aggregate impacts of foreigncompetition on employment.
27
This figure comes from information provided in table 2 of Bernard et al.
ð2006Þ.

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centage points ðtable 1Þ. Had their much lower level of import growthobtained over our sample period, the reduction in manufacturing job loss
implied by our coefficient estimates would have been only one-fifth aslarge.28 One reason why Bernard et al.’s analysis may produce higher estimates of the impact of imports on employment than ours is that they
study plant-level data as compared to our industry-level regressions. Aggregating across plants within an industry is preferable in this instance because it avoids confounding aggregate effects with within-industry reallocation, which take place as some workers may exit declining plants to takejobs with establishments in their same sector ðconsistent with the results inAutor et al. ½2014Þ.
Pierce and Schott ð2015Þ use a difference-in-difference strategy to test
whether after 2001 manufacturing employment fell by more in industries
that were more exposed to China’s WTO accession. They measure thispotential increase in exposure to China trade using the difference between
the US MFN ðmost-favored-nationÞ tariff and the US non-MFN tariff, towhich China was potentially subject prior to becoming a WTO member
and whose level was substantially higher than the MFN duty. Pierce andSchott thus identify the growth in China trade after 2001 using the notional reduction in US trade barriers confronting China. A complication
with this approach is that the United States granted China MFN status
on a renewable basis in 1980, 2 decades prior to the country joining the
WTO. The US non-MFN tariff is a meaningful predictor of China’s pre2001 trade only to the extent that there was genuine risk the US government would choose not to renew China’s MFN privileges, an eventualitythat Congress discussed annually but that never materialized. Pierce andSchott estimate that China’s WTO accession reduced post-2001 manufacturing employment by 15.1 log points in exposed industries relative tononexposed industries.29 Our estimates, which identify the impact of growthin China’s imports based on the common component of the country’s export expansion across high-income markets, imply that had there been no
increase in import penetration from China after 1999, the 2011 level of
employment would have been 4.9% higher ð0.560 million/11.4 millionÞ
than it otherwise would have been. Comparing our results in table 2 tothose of Bernard et al. ð2006Þ and Pierce and Schott ð2015Þ thus suggeststhat our estimates for the direct industry-level employment effects of China
trade are relatively modest.
28
This ratio is based on the calculation ð1 2 e21:30:56:09 Þ=ð1 2 e21:30:56:50 Þ 5
0:21, where the value 21.30 is the coefficient from col. 3 of table 2 and the value.56 discounts observed changes in import penetration by the partial R-squared of
the first stage.
29
This estimate is from col. 3 of table 1 of their paper, which we view as closest
in spirit to the specifications in our article.

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C. Controlling for Industry Confounds and Pretrends
A challenge for our analysis is that industries subject to greater importcompetition may be exposed to other economic shocks that are correlatedwith China trade. We begin to address this concern in table 3 by incorporating controls for potential industry confounds. We additionally offer
a set of falsification tests.
We consider three groups of control variables. First, we probe the robustness of our results by including dummies for 10 one-digit manufacturing sectors. Since our regressions are in first differences, the inclusionof these dummies amounts to allowing for differential trends across theseone-digit sectors. Regressions including these dummies therefore identify
the industry-level impacts of trade exposure while purging common trends
within the one-digit sectors and using only variation in import growthacross industries with relatively similar skill intensities.
Technological progress within manufacturing has been most rapid inrecent decades in computer and skill-intensive sectors ðDoms, Dunne, andTroske 1997; Autor, Katz, and Krueger 1998Þ. To capture the extent towhich industries are exposed to technical change, we next add a second setof control variables, drawn from the NBER-CES database, measuring theintensity of their use of production labor and capital. These variables,
summarized in table A1, include the share of production workers in total
employment, the log of the average wage, the ratio of capital to value
added ðall measured in 1991Þ, as well as computer and high-tech equipment investment in 1990, each expressed as a share of total 1990 investment.
US manufacturing as a share of employment has been declining sincethe 1950s, and the number of manufacturing employees has also trended
downward since the 1980s. This long-standing secular trend highlights aconcern that the correlation we document between rising industry trade
penetration and contemporaneous, within-industry declines in manufacturing employment during 1991–2011 could potentially predate the recent
rise in import exposure. In that case, our estimates would likely overstate
the impact of trade exposure in the current period. We therefore finally
add measures of pretrends in industry employment and earnings in table 3,
specifically the change in the industry’s share of total US employment andthe change in the log of the industry average wage, both measured over theinterval 1976–91 ðtable A1Þ.
The first seven columns of table 3 permute among combinations of thesethree groups of industry controls: the one-digit sector dummies, industrylevel controls for production structure, and industry-level controls for
pretrends. Column 1 replicates results from column 3 of table 2 to serve as abenchmark. Among the additional groups of covariates, only the one-digit
sector dummies have a substantial impact on the point estimates, reducing

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21.30***
ð.41Þ
.05
ð.36Þ
23.46***
ð.33Þ
No
No
No
No

2.75***
ð.22Þ
2.09
ð.32Þ
23.82***
ð.27Þ
Yes
No
No
No

ð2Þ
21.10***
ð.35Þ
.00
ð.37Þ
23.59***
ð.35Þ
No
Yes
No
No

ð3Þ
21.33***
ð.43Þ
.06
ð.36Þ
23.44***
ð.32Þ
No
No
Yes
No

ð4Þ
2.80***
ð.25Þ
2.08
ð.30Þ
23.79***
ð.28Þ
Yes
Yes
No
No

ð5Þ
2.76***
ð.22Þ
2.09
ð.32Þ
23.82***
ð.26Þ
Yes
No
Yes
No

ð6Þ

2.74***
ð.23Þ
2.10
ð.30Þ
23.83***
ð.27Þ
Yes
Yes
Yes
No

ð7Þ

23.79***
ð.45Þ
No
No
No
Yes

2.60**
ð.29Þ

ð8Þ

NOTE.—Each column reports results from stacking log employment changes and changes in US exposure to Chinese imports over the periods 1991–99 and 1999–2011 ðN 5 784 5
392 four-digit manufacturing industries  2 periodsÞ. The dependent variable is 100  the annual log change in each industry’s employment in the CBP over the relevant period. The
regressor is 100  the annual change in US exposure to Chinese imports over the same period; it is instrumented as described in the text. Sector controls are dummies for 10 one-digit
manufacturing sectors. Production controls for each industry include production workers as a share of total employment, the log average wage, and the ratio of capital to value
added ðin 1991Þ; and computer investment as a share of total investment and high-tech equipment as a share of total investment ðin 1990Þ. Pretrend controls are changes in the log
average wage and in the industry’s share of total employment over 1976–91. In the final column, we include a full set of four-digit industry fixed effects. Covariates are demeaned tofacilitate interpretation of the time effects. Observations are weighted by 1991 employment. Standard errors in parentheses are clustered on 135 three-digit industries.
** p < .05.
*** p < .01.

One-digit manufacturing sector controls
Production controls
Pretrend controls
Industry fixed effects

1f1999–2011g

1f1991–99g

100  annual D in US exposure
to Chinese imports

ð1Þ

Table 3
2SLS Estimates of Import Effects on Log Employment Including Industry-Level Controls

Import Competition and the Great US Employment Sag

S165

the ðinstrumentedÞ estimates by about 40%.30 Though the inclusion of thesectoral dummies is an important robustness check for our results, there
are two reasons why these specifications may underestimate the impactof Chinese import competition. First, trade exposure at the four-digit
industry level is likely to be measured with error, and the inclusion of the
one-digit sector dummies will then cause significantly greater attenuationof our estimates of the impact of Chinese import growth. Second, if there isa significant increase in imports in some industries within a one-digit sector
ðsay, in women’s dresses within textilesÞ, then employers in other similarindustries within this broad sector ðsay, women’s blouses and shirts, also
within textilesÞ may anticipate greater competition both from the substitutes already being imported from China and also from future waves ofChinese imports and thus will be more likely to downsize and close existing
plants and less likely to open new plants. By contrast, neither the production nor the pretrend variables have an important effect on the magnitude or precision of the coefficient of interest. As a further robustness
test, column 8 includes a full set of dummies for the 392 four-digit manufacturing industries in our data. These variables serve as industry-specifictrends in our stacked first-difference specification, so the effect of importcompetition on industry employment in this specification is identified bychanges in the growth rates of industry employment and import penetration in 1999–2011 relative to 1991–99. Remarkably, relative to specifications that include one-digit sector dummies, the addition of an exhaustive
set of industry-specific trends only modestly reduces the point estimate andprecision of the coefficient of interest, thus highlighting the robustness of
the relationship. In summary, while our preferred industry-level modelfrom column 3 of table 2 allows for an impact of Chinese trade competition
on employment both within and across broad manufacturing subsectors,
the estimates in table 3 document that a sizable negative employment effect remains even when focusing only on the within-subsector or withinindustry, over-time variation in trade exposure.
As a falsification exercise, table 4 reports results from a regression ofchanges in industry employment in earlier decades on the instrumentedchange in industry import exposure between 1991 and 2011. It would be
problematic for our identification strategy if future growth in Chinese
import exposure predicted industry employment declines in the era prior

30

Quantitatively, the specification in col. 2 of table 3 implies that had importpenetration from China remained unchanged between 1991 and 2011, manufacturing employment would have fallen by 463,000 jobs over the full 1991–2011 spanand by 307,000 jobs between 1999 and 2011; these figures are about 45% lower
than our baseline numbers.

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S166

Acemoglu et al.

Table 4
2SLS Estimates of Import Effects on Log Employment over 1971––2009
1971–81ð1Þ

1981–91
ð2Þ

1991–99
ð3Þ

1999–2009ð4Þ

1991–2009
ð5Þ

A. Excluding One-Digit Manufacturing Sector Controls
100  annual D in US
exposure to Chinese importsðcomputed over 1991–2011Þ
Constant.34
ð.33Þ
1.19***
ð.30Þ

2.40
ð.28Þ
2.68**
ð.34Þ

2.84*
ð.45Þ
.35
ð.46Þ

22.01***
ð.66Þ
23.97***
ð.43Þ

21.49***
ð.51Þ
22.05***
ð.29Þ

B. Including One-Digit Manufacturing Sector Controls
100  annual D in US
exposure to Chinese importsðcomputed over 1991–2011Þ
Constant

.20
ð.26Þ
2.05
ð.32Þ

.03
ð.26Þ
2.08
ð.74Þ

2.57*
ð.31Þ
.52
ð.63Þ

2.91***
ð.31Þ
2.98**
ð.45Þ

2.76***
ð.23Þ
2.32
ð.48Þ

NOTE.—N 5 384 four-digit manufacturing industries ðwe exclude eight industries for which post-1996
employment data are unavailable in the NBER-CES Manufacturing Industry DatabaseÞ. The dependent
variable in each specification is 100  the annual log employment change over the indicated period, as
computed in the NBER-CES data. The regressor in each specification is 100  the annual change in US
exposure to Chinese imports over 1991–2011, instrumented as described in the text. Panel A includes noadditional controls. Panel B includes dummies for 10 one-digit manufacturing sectors. Observations areweighted by 1991 employment. Standard errors in parentheses are clustered on 135 three-digit industries.
* p < .10.
** p < .05.
*** p < .01.

to China’s trade opening.31 Panel A performs this exercise without additional covariates, while panel B controls for 10 one-digit sector dummies.
In both panels, the estimated relationship between our China trade exposure measure and industry employment is statistically insignificant andclose to zero in both the 1970s ð1971–81Þ and 1980s ð1981–91Þ. The point
estimate becomes economically large and statistically significant only after
1990. This pattern of results is consistent with the hypothesis that thewithin-industry correlation between rising import penetration and declining
manufacturing employment in the 1990s and 2000s emanates from con-31
To carry the analysis back to 1971, we employ the NBER-CES data, which
cover a longer time horizon than the CBP data used in our main estimates. A
disadvantage is that the NBER-CES database is currently updated only through2009, 2 years less than the CBP. To improve comparability, we use the NBER datain all columns of table 4, including for the post-1990 period ðin contrast to tables 2
and 3, where we use CBP dataÞ. These estimates also differ from those in tables 2
and 3 in that the import exposure variable ðand its instrumentÞ corresponds to thelong 1991–2011 change in all columns.

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Import Competition and the Great US Employment Sag

S167

temporaneous trade shocks rather than from long-standing factors driving industry decline.
D. Additional Employment and Establishment-Level Outcomes
We have so far focused on the effects of trade exposure on industry employment, which is but one margin along which industries adjust. Others
include the wage bill, establishment size, establishment shutdown, and production versus nonproduction employment and earnings. Using a combination of CBP and NBER-CES data, we explore these outcomes in table 5.
Given our findings on how import penetration affects employment in
tables 2 and 3, many of the results in table 5 are in line with expectations.
Stronger import competition reduces the count of establishments ðcol. 2Þ,average employment per establishment ðcol. 3Þ, and total industry wage
payments ðcol. 4Þ. Production employment ðcol. 6Þ declines slightly morethan nonproduction employment ðcol. 7Þ, indicating a larger sensitivity to
Chinese import competition on the part of lower-skilled labor, a result
consistent with China’s strong comparative advantage in labor-intensive
sectors.
The table also contains some informative surprises. Trade exposure predicts a rise in real industry log wages for production workers ðcol. 8Þ—that
is, the real production worker wage bill divided by the production worker
headcount. The impact on nonproduction worker wages ðcol. 9Þ is negative but small and not statistically significant. Joining these two effects
produces the positive but insignificant coefficient estimate for average real
wages ðcol. 5Þ. The results for production workers that combine strongly
negative employment effects and mildly positive average wage effects aresuggestive of trade-induced changes in the composition of employment.
Less highly paid workers may be those more likely to be laid off within thesubgroup of production employees, leading to an upward shift in wages
among those still employed as a result of unobserved changes in composition. This interpretation is consistent with Autor et al.’s ð2014Þ finding thatthe earnings of lower-wage workers are most adversely affected by greater
import competition.32
32
Complementing these results, table A2 reports the impact of Chinese importcompetition on industry output, measured as the value of shipments. In panel A,
we find that import exposure has an economically and statistically significant negative effect on nominal shipments ðcol. 1Þ; but when we decompose this effectinto changes in real shipments and changes in the shipments price deflator ðcols. 2
and 3Þ, we find no effect on real shipments. This surprising pattern turns out to be
driven by computer-producing industries, which experienced rapid growth in realvalue added, precipitous declines in output prices, and substantial increases in
Chinese import penetration during our sample period. In panel B, where we exclude
28 computer-producing industries corresponding to North American Industry

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21.67***
ð.17Þ

.41**
ð.19Þ

2.30
ð.32Þ

24.32***
ð.25Þ

Yes

21.51***
ð.19Þ

2.23***
ð.09Þ
.48**
ð.19Þ

Yes

23.82***
ð.27Þ

2.75***
ð.22Þ
2.09
ð.32Þ

22.66***
ð.17Þ

2.71***
ð.26Þ

Yes

22.31***
ð.18Þ

2.52***
ð.17Þ
2.57**
ð.26Þ
23.42***
ð.30Þ

2.67***
ð.21Þ
1.53***
ð.30Þ

Yes.40***
ð.10Þ

.08
ð.06Þ
1.63***
ð.08Þ

23.87***
ð.29Þ

1.35***
ð.31Þ
.46***
ð.08Þ

1.65***ð.07Þ

NBER-CES DatabaseYes

24.84***
ð.37Þ

2.99***
ð.31Þ
.33
ð.38Þ

25.38***
ð.34Þ

.06
ð.38Þ

24.06***
ð.32Þ

2.40
ð.33Þ

Yes

23.63***
ð.31Þ

2.78***
ð.29Þ
2.20
ð.34Þ

.35**
ð.14Þ

1.19***
ð.06Þ

Yes

.22
ð.14Þ

.24**
ð.11Þ
1.13***
ð.06Þ

.30**
ð.13Þ

1.80***
ð.08Þ

Yes.32***
ð.11Þ

2.05
ð.09Þ
1.81***
ð.09Þ

Real
Real
Production Nonproduction Production Nonproduction
Wage
Employment Employment
Wage
ð6Þ
ð7Þ
ð8Þ
ð9Þ

B. Dependent Variable Means by Time Period

Yes

Real
Wage
ð5Þ

A. 2SLS Estimates

Real
Wage
Bill
ð4Þ

NOTE.—Each column stacks changes in the indicated outcome and changes in US exposure to Chinese imports over the periods 1991–99 and either 1999–2011 ðfor CBP outcomesÞ
or 1999–2009 ðfor NBER-CES outcomesÞ. In cols. 1–5, N 5 784 5 392 four-digit manufacturing industries  two periods. In cols. 6–9, we exclude eight industries for which post1996 data are unavailable in the NBER-CES, yielding N 5 768 5 384 industries  two periods. In each column, the dependent variable is 100  the annual log change in the indicatedquantity. Panel A reports 2SLS estimates including the annual change in US exposure to Chinese imports over the relevant period; it is instrumented as described in the text. Panel B
reports OLS estimates from a regression including only time effects and sector controls. All specifications include dummies for 10 one-digit manufacturing sectors, which aredemeaned to facilitate interpretation of the time effects. Observations are weighted by 1991 employment in the relevant data set. Standard errors in parentheses are clustered on 135
three-digit industries.
** p < .05.
*** p < .01.

1f1999–2011g or
1f1999–2009g

1f1991–99g

One-digit manufacturingsector controls

1f1999–2011g or
1f1999–2009g

1f1991–99g

100 annual D in
US exposure to
Chinese imports

N
Employment
Employment Establishments per Establishment
ð1Þ
ð2Þ
ð3Þ

County Business Patterns

Table 5
2SLS Estimates of Import Effects on Additional Labor Market OutcomesImport Competition and the Great US Employment Sag

S169

V. Accounting for Sectoral Linkages
We now expand the scope of the inquiry to encompass the effects oftrade shocks on employment in both manufacturing and nonmanufacturing industries working through input-output linkages. In appendix B,
we present a simple model of Cobb-Douglas production that yields expressions for changes in industry employment resulting from upstream
and downstream import exposure. Here we discuss the empirical implementation of these upstream and downstream effects.
To study these interindustry linkages, we envisage an economy along
the lines of that studied by Long and Plosser ð1983Þ and Acemoglu et al.
ð2012Þ, where each industry uses with different intensities the output ofother industries as inputs. We apply this methodology to the BEA’s inputoutput table for 1992. We choose the 1992 input-output table since itlargely predates the China trade shock and hence measures linkages thatare unlikely to be endogenous to the subsequent shock.
To estimate the upstream effect—the exposure to import competition
that propagates upstream from an industry’s buyers—we calculate thefollowing quantity for each industry j:
DIPUjt 5 o wUgj DIPgt ;ð5Þgwhich is equal to the weighted average change in import penetration during time interval t across all industries, indexed by g, that purchase fromindustry j. These weights wDgj are defined as
wDgj 5

mUgj

om
g0

U
g0 j

;

ð6Þ

where mUgj is the 1992 “use” value in the BEA input-output matrix for thevalue of industry j’s output purchased by industry g, such that the weightin ð6Þ is the share of industry j’s total sales that are used as inputs byindustry g. Thus, ð5Þ is a weighted average of the trade shocks faced by thepurchasers of j’s output.33 When industry j’s purchasers suffer a negative
Classification System ðNAICSÞ 334, we find comparable effects on nominal shipments, but these effects are now driven primarily by relative declines in real shipments in trade-exposed industries rather than by relative declines in output prices.
We view these results as consistent with a mounting body of evidence that computerproducing industries have an outsized influence on measured output and productivity
in the manufacturing sector ðAcemoglu et al. 2014b; Houseman, Bartik, and Sturgeon
2015Þ.
33
We use the BEA “make” table to assign commodities to the industries that
produce them. The summation in the denominator of eq. ð6Þ runs over not only
manufacturing industries but also nonmanufacturing industries as well as final
demand. Since our direct shock variable reflects only manufacturing trade, allThis content downloaded from 23.235.32.0 on Fri, 18 Dec 2015 07:23:05 AM
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S170

Acemoglu et al.

trade shock, they are likely to reduce demand for j’s output. The theoretical justification for these expressions is provided in appendix B using asimple model of input-output linkages.
Similarly, to compute the downstream effect DIPDjt experienced by eachindustry j—that is, the exposure to import competition that propagatesdownstream from j’s suppliers—we make the same calculation after reversing the j and g indexes in the numerator of ð6Þ.34 We instrument boththe upstream and downstream exposure measures analogously to our main
import shock measure: using contemporaneous changes in China imports
in eight other high-income countries to calculate predicted upstream anddownstream exposure for each industry, where these predictions serve as
instruments for the measured domestic values. Concretely, we constructthese instruments by replacing the term DIPgt with DIPOgt in equation ð5Þwhile retaining the same weights.
Equation ð5Þ accounts for the direct ðfirst-orderÞ effect on output demand of an industry j stemming from trade-induced changes in demand
from its immediate buyers. But it ignores further indirect effects on industry j’s demand stemming from changes in demand from its buyers’ buyers, and so on. To account for the full chain of linked downstream and upstream demands, we replace DIPUjt and DIPDjt ðand their instrumentsÞ withthe full chain of implied responses from the input-output matrix, which is
given by the Leontief inverse of the matrix of upstream and downstream
linkages ðsee, e.g., Acemoglu et al. 2012Þ. The details of this computation
are given in appendix B.
Upstream and downstream exposure measures are summarized in table A3. As expected, the indirect exposure measures are substantiallysmaller in magnitude, and have far less cross-industry variation, than the
direct exposure measures. In the average manufacturing industry, direct
trade exposure is five times as large as the first-order downstream exposure
measure and over three times as large as the first-order upstream exposure measure. Incorporating higher-order linkages significantly increases
the magnitude of the upstream and downstream exposure measures. The
full indirect upstream exposure measure ðgiven by the Leontief inverseÞ isapproximately half as large as the direct exposure measure, while the full
upstream effects experienced by a sector emanate by definition from shocks totheir manufacturing purchasers ði.e., DIPgt is defined to equal zero for nonmanufacturing industries and for final demandÞ. These shocks affect both manufacturing and nonmanufacturing industries to the degree that they supply inputs to
manufacturing industries g that are directly shocked.
34
When we construct weights for the downstream effect, the summation in the
denominator again runs over industry j’s total sales. Analogously to the case of
upstream effects, downstream effects emanate from trade shocks to these industries’ suppliers in manufacturing ðthough, as just noted, both manufacturers andnonmanufacturers may have suppliers in manufacturingÞ.

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Import Competition and the Great US Employment Sag

S171

indirect downstream exposure measure is about one-third as large as the
direct exposure measure.
The two panels of table 6 present instrumental variables estimates ofthe effects of import exposure on industry employment, akin to those intable 3, column 1 ðwithout the one-digit sector dummiesÞ and column 2ðwith the one-digit sector dummiesÞ, here augmented with the upstream
and downstream import exposure measures. Panel A of table 6 employs
the first-order upstream and downstream measures, DIPUjt and DIPDjt , whilepanel B uses the full Leontief exposure measures. We present results withand without the one-digit sector dummies introduced earlier.35
Columns 1–3 of table 6 consider the impact of upstream and downstream linkages on employment in the 392 manufacturing industries; columns 4 and 5 consider these impacts on employment in the 87 nonmanufacturing industries; and columns 6–10 present results for manufacturing
and nonmanufacturing pooled. All regressions employ the stacked firstdifferences specification: columns 1–8 and 10 cover the time periods 1991–
99 and 1999–2011, while column 9 shortens the second period to 1999–
2007. Downstream import effects are not statistically significant in any
specification and are unstable in sign, showing up as positive in the manufacturing only specification ðcol. 2Þ and negative in the nonmanufacturing
and pooled specifications ðcols. 5 and 7Þ.36 This imprecision may be dueto the fact that the downstream effects combine the offsetting effects ofreduced domestic input supply ðdue to US-based suppliers curtailing shipments in the face of increased import competitionÞ and increased foreign
input supply. Given the instability of effects working through downstream
linkages, we focus our attention on the upstream effects, which are, incontrast, quite stable across specifications and are qualitatively similar for
manufacturing and nonmanufacturing sectors.
Consistent with our reasoning above, growth in an industry’s upstream
trade exposure is found to reduce industry employment. For manufacturing industries alone, the coefficient of the upstream linkage effect isquite large without the one-digit sector dummies in the regression ðcol. 2Þ
and has a magnitude similar to that of the direct trade shock coefficient aswell as more precisely estimated when the one-digit sector dummies areadded in column 3. For nonmanufacturing industries, upstream linkages
are also negative and statistically significant ðcols. 4 and 5Þ and larger in
35
We do not include the industry production and pretrend controls used intable 3. These were shown to have little effect conditional on sector dummies but
still absorb degrees of freedom, which is problematic in a setting with multiple instrumented endogenous variables that are themselves correlated.
36
Additionally, the downstream effect in manufacturing reverses sign ðwhileremaining insignificantÞ when the upstream variable is omitted. Observe that there
is no “direct” trade exposure effect in nonmanufacturing since our trade measures
are confined to manufactured goods.

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Yes
No
No

No
No

21.30***
ð.49Þ
21.78**
ð.82Þ
1.74
ð2.10Þ

21.20***
ð.42Þ
21.64*
ð.84Þ

Yes

21.28***
ð.49Þ
22.44**
ð1.13Þ
2.31
ð2.66Þ

21.17***
ð.42Þ
22.21*
ð1.14Þ

Yes
No

Yes

2.72***
ð.22Þ
2.85**
ð.37Þ

2.72***
ð.22Þ
21.03**
ð.45Þ

ð3Þ

ð5Þ

ð6Þ

ð7Þ

26.88**
ð2.97Þ
25.80
ð7.43Þ

21.14***
ð.42Þ
22.70**
ð1.26Þ

21.11**
ð.48Þ
22.64**
ð1.32Þ
2.67
ð3.69Þ

No
No

Yes

23.19
ð2.14Þ

No
No

Yes

23.17
ð2.27Þ
24.26
ð5.94Þ

No
No

Yes

21.18***
ð.42Þ
21.90**
ð.86Þ

No
No

Yes

21.14**
ð.48Þ
21.86**
ð.91Þ
2.68
ð2.95Þ

B. Full ðHigher-OrderÞ Input-Output Linkages

26.63**
ð2.79Þ

Yes
No

Yes

2.71***
ð.22Þ
21.29**
ð.59Þ

2.69***
ð.22Þ
21.72**
ð.75Þ

ð8Þ

No
Yes

Yes

21.12***
ð.38Þ
22.10***
ð.75Þ

21.07***
ð.38Þ
23.06***
ð1.09Þ

ð9Þ

Pooling Manufacturing andNonmanufacturing Industries
ðN 5 958Þ

A. First-Order Input-Output Linkages

ð4Þ

Nonmanufacturing
Industries
ðN 5 174Þ

NoNo

21.32***
ð.37Þ
Yes

1.35***
ð.38Þ

ð10Þ

NOTE.—The sample consists of 392 manufacturing industries ðcols. 1–3Þ, 87 nonmanufacturing industries ð4–5Þ, or both sets of industries pooled ð6–10Þ. Each column stacks
changes in log employment and changes in import exposure over the periods 1991–99 and either 1999–2011 ðcols. 1–8, 10Þ or 1999–2007 ð9Þ. The dependent variable is 100  theannual log change in employment, as computed in the CBP. The direct import exposure of industry i equals 100  the annual change in US exposure to Chinese imports. In panel A,
upstream ðrespectively, downstreamÞ import exposure for a given industry is a weighted average of the direct import exposure experienced by its customers ðsuppliersÞ, as identified
by the Bureau of Economic Analysis’s 1992 input-output table. In panel B, we use the Leontief inverse of the input-output matrix to incorporate higher-order linkages. Direct,
upstream, and downstream measures of import exposure are instrumented using changes in comparison countries’ exposure to Chinese imports. See the text for details. In col. 10,
combined import exposure is defined as the sum of the direct and upstream exposure measures used in the other columns; we include separate instruments for the direct andupstream components of the combined measures. Columns 1–5 include dummies for each time period. Columns 6–10 include sector ðmanufacturing/nonmanufacturingÞ  periodinteractions. Where indicated, we include dummies for 10 one-digit manufacturing sectors ðwhich equal zero for nonmanufacturing industriesÞ. Observations are weighted by 1991
industry employment, and standard errors in parentheses are clustered on three-digit industry ðwith each nonmanufacturing industry constituting its own clusterÞ.
* p < .10.
** p < .05.
*** p < .01.

Sector  period effects
One-digit manufacturing sector
controls
Exclude 2007–11

Combined import exposureðdirect 1 upstreamÞ

Downstream import exposure

Upstream import exposure

Direct import exposure

Combined import exposureðdirect 1 upstreamÞ

Downstream import exposure

Upstream import exposure

Direct import exposure

ð2Þ

ð1Þ

Manufacturing Industries
ðN 5 784Þ

Table 6
2SLS Estimates of Import Effects on Employment Incorporating Input-Output Linkages

Import Competition and the Great US Employment Sag

S173

magnitude than the estimates for manufacturing. Pooling manufacturing
and nonmanufacturing, coefficients on upstream linkages are negative andstatistically significant either without ðcols. 6 and 7Þ or with ðcol. 8Þ the onedigit sector dummies included in the regression.37 Results for the period
1991–2007 ðcol. 9Þ are quantitatively similar.
Finally, in the last specification in panel B ðcol. 10Þ, we regress changes
in industry employment on the sum of the direct and upstream exposure
measures, which is the form suggested by our theoretical model in appendix B. As expected, the estimated coefficient on the combined shock lies
between the coefficients on the direct and upstream effects in column 6.38
Comparing across the two panels of table 6, which employ the first-orderðpanel AÞ and full ðpanel BÞ upstream and downstream measures, we detect
a similar pattern of coefficient estimates. In all cases, the coefficients on thefull exposure measures are smaller in magnitude than those on the firstorder exposure measures, though they are also more precisely estimated.
Of course, the full exposure measures are considerably larger in magnitude than the first-order exposure measures, so the smaller coefficients do
not imply smaller quantitative effects.
Accounting for upstream linkages substantially increases the impact oftrade shocks on employment. Using estimates from the regression thatpools manufacturing and nonmanufacturing together ðcol. 6, the specification without one-digit sector dummiesÞ, we evaluate the counterfactualchange in employment analogous to the exercise in equation ð4Þ, with theresults again shown in table 8. This new exercise combines the employment impacts of trade shocks working through direct effects and indirect
effects associated with upstream linkages.39 Had import competition from
China remained unchanged between 1991 and 2011, according to our estimates from panel A ðusing only first-order upstream effectsÞ, there would
have been 1.33 million additional workers employed in manufacturing and805,000 additional workers employed in nonmanufacturing, for a total employment differential of 2.14 million workers. Examining just the 1999–2011
period, the corresponding counterfactual employment additions are 928,000
The nonmanufacturing estimates do not include sector dummies ðunlike the
manufacturing estimatesÞ since our nonmanufacturing industry scheme is alreadyhighly aggregated and, moreover, does not collapse down readily to a one- or twodigit sector scheme since we had to extensively aggregate four-digit SIC industries
for concordance with the input-output tables used by the BEA.
38
We cannot reject the hypothesis that the coefficient on this combined variable
is the same as the separate coefficients on the direct and the upstream exposure
measures in col. 2. The implied quantitative magnitudes ðreported belowÞ are also
very similar regardless of whether we use this combined measure or separate measures for direct and indirect upstream effects.
39
Consistent with the analysis of Sec. IV, these counterfactuals assume that 56%of the observed growth in direct and indirect import exposure is attributable to theChinese supply shock.
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S174

Acemoglu et al.

in manufacturing and 653,000 in nonmanufacturing, for a total of 1.58 million additional workers employed. Accounting for the full set of direct andindirect upstream effects shown in our preferred specification ðpanel B,
col. 6Þ, we obtain employment estimates that are larger again: 1.41 million
workers in manufacturing, 1.22 million in nonmanufacturing, and 2.62 million overall for 1991–2011; and 985,000 workers in manufacturing, 994,000
in nonmanufacturing, and 1.98 million overall for 1999–2011. These combined direct and indirect effects of increased Chinese imports are substantially larger than the direct effects alone ð837,000 workers for 1991–2011 and
560,000 workers for 1999–2011Þ. Thus, accounting for upstream linkages
inside and outside of manufacturing more than triples the estimated direct
employment effects for manufacturing alone.40
These estimated magnitudes do not, however, include the full generalequilibrium impact of trade exposure as they fail to capture aggregate reallocation and demand effects as outlined above. We turn to local labor market
analysis to obtain estimates of these additional adjustment mechanisms.
VI. Local General Equilibrium Effects of Trade on Employment
Our industry-level analysis, which compares changes in relative employment among industries with differing levels of trade exposure, is not well
suited to identifying the reallocation and demand effects discussed in the
introduction and Section II. In this section, we attempt to quantify the reallocation and aggregate demand effects by applying an alternative strategy
that focuses on the implications of rising import competition from China
for employment in local labor markets.
A. Empirical Approach
To exposit the logic of our approach, consider a simplified setting in which
each commuting zone ðCZÞ houses up to three sectors that have no inputoutput linkages: toys, footwear, and construction.41 Toys and footwear
40

The specification in col. 8, which controls for 10 one-digit manufacturingsector dummies, implies somewhat smaller employment effects. According to ourestimates from panel B ðaccounting for the full set of direct and upstream effectsÞ,had import competition from China remained unchanged between 1991 and 2011,
there would have been 857,000 additional workers employed in manufacturing
and 821,000 additional workers employed outside of manufacturing, for a total
employment gain of 1.68 million workers. For the 1999–2011 period, the corresponding counterfactual employment additions are 597,000 in manufacturing and
670,000 in nonmanufacturing, yielding total employment gains of 1.27 million.
These numbers are about 35% smaller than our baseline estimates incorporating
the indirect upstream effects.
41
The choice of construction as the nontraded sector is motivated in part by thestudy by Charles, Hurst, and Notowidigdo ð2013Þ, who find that the 2000–2007
housing boom helped local labor markets absorb workers displaced from manufacturing.

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Import Competition and the Great US Employment Sag

S175

experience an increase in imports from China, so we label these sectors as
exposed. Construction does not experience this shock, and we label it nonexposed. If a particular CZ has many workers employed in toys prior to
the rise of import competition from China, it will experience significant
worker displacement as this sector contracts.42 Because of the reallocation
effect, we would expect displaced workers to gain employment in anothersector. This sector is unlikely to be footwear, however, since it is simultaneously facing rising import competition. In this simple setting, labor within
the CZ should therefore reallocate toward construction. Estimating by how
much employment in construction expands in this CZ as toys and footwear
decline can help us to assess the positive general equilibrium effects resulting
from reallocation.
Employment in construction may be affected by a second channel as well:
the potentially negative Keynesian aggregate demand multiplier, stemmingfrom reductions in local economic activity. In our simple example, the initial
reduction in employment in exposed industries will reduce local incomes
and, via this channel, may depress local demand for new home construction
or renovation, further depressing employment.43 The net effect of these reallocation and aggregate demand effects on employment in construction may
be positive or negative.
Now suppose that the third industry in this economy is not construction
but chemicals, which, unlike construction, is tradable within the UnitedStates across local labor markets and, as it happens, has not been subject tosignificant increases in import competition from China. To make progressin this case, suppose that our local labor markets can be thought of as small
open economies within the United States, so that prices of tradables are
determined at the US level ðor on world marketsÞ. This does not change thereallocation effect, but it may alter the aggregate demand effect. Even if
aggregate demand for nontradables in the local labor market is depressed,
there might be an increase in local employment in chemicals, the output of
which is then sold to residents in other CZs. This is simply a reflectionof the fact that the component of the negative aggregate demand effectworking at the national level will not be easily identified from variation
across local labor markets. An implication of this observation is that our
strategy will tend to underestimate the aggregate demand effect to thedegree it operates nationally rather than locally.

42
This discussion also makes it clear that empirically it is appropriate tocombine the shocks of all of the local industries using weights related to theirlocal employment shares, which is the strategy employed here and in Autor et al.
ð2013Þ.
43
It is possible for trade-induced price declines to simultaneously contribute toaggregate demand by spurring additional consumption or investment, as discussed
in footnote 14.

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B. Estimates
The local labor market analysis is based on 722 CZs that cover the entire
US mainland. These CZs are clusters of counties with strong internal
commuting ties ðsee Tolbert and Sizer 1996; Autor and Dorn 2013Þ.
We begin by estimating stacked first-difference models for changes in
CZ employment-to-population rates of the following form:
DEit 5 at 1 bDIPCZ
it 1 gXi0 1 eit :ð7Þ

Here, the dependent variable DEit is equal to 100 times the annual change
in the ratio of employment to working-age population in CZ i over time
period t; Xi0 is a set of CZ-by-sector start-of-period controls ðspecified
laterÞ; at is a time effect; and eit is an error term.44 The key explanatory
variable in this model is DIPCZ
it , which measures a CZ’s annual change in
exposure to Chinese imports over period t. The coefficient b reveals theimpact of import exposure on overall employment rates, combining employment shifts in both trade-exposed and nonexposed industries. We
define a CZ’s change in import exposure as a local employment-weighted
average of changes in import exposure:
DIPCZ
it 5 oj

Lijt
DIPjt :
Litð8ÞIn ð8Þ, DIPjt is the measure of Chinese import competition used in our
industry-level analysis, and Lijt/Lit is industry j’s start-of-period share of
total employment in CZ i.45 The variation in DIPCZ
it across local labor markets stems entirely from variation in local industry employment structure
at the start of period t. As with our industry-level estimates, a concern isthat realized US imports from China in ð8Þ may be correlated with industry
import demand shocks. We again instrument for growth in Chinese imports
to the United States using the contemporaneous growth of Chinese imports
in eight other developed countries as specified in ð2Þ.46 Table A4 summarizes
44

Throughout this section, local employment is derived from the CBP, andlocal working-age population ðages 15–64Þ is derived from the Census of Population estimates.
45
This is similar to Autor et al. ð2013, 2014Þ, except that for consistency with
our industry-level analysis, we normalize industry-level imports by initial US
market volume instead of initial employment.
46
Our expression for non-US exposure to Chinese imports, which serves as an
instrument for DIPCZ
it , differs from the expression in eq. ð8Þ in that in place ofrealized changes in US import exposure ðDIPjtÞ, we use the analogous expression
based on realized imports from China to other high-income markets ðDIPOjtÞ.
In addition, we use 1988 employment counts for the construction of the instrument to reduce the error covariance between the dependent and independent
variables.

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Import Competition and the Great US Employment Sag

S177

CZ-level changes in exposure to Chinese imports and in employment-topopulation rates.
To gauge the differential impact of import exposure on different types ofindustries within local labor markets, we decompose employment changes
into three broad sectoral groupings. Specifically, we interact the CZ’s change
in import exposure with indicator variables for exposed industries, nonexposed tradable industries, and other nonexposed industries:
 1½Exposedk 
DEikt 5 akt 1 b1 DIPCZ
it 1½Nonexposed Tradablek 
1 b2 DIPCZ
it ð1 2 1½Exposedk 
1 b3 DIPCZ
itð9Þ

2 1½Nonexposed Tradablek Þ 1 gXik0 1 eikt :In these regressions, DEikt is the change in employment of sector k in CZ i,
expressed in percentage points of working-age population. While thespecification in ð9Þ is similar to that in Autor et al. ð2013Þ, it differs importantly by separating the employment effects of import competition in
CZs according to sector import exposure and tradability. To compute DEikt,
we assign each industry to one of the three mutually exclusive sectors:
exposed industries, nonexposed tradable industries, and other nonexposedindustries. First, we define the exposed sector to encompass all manufacturing industries for which predicted import exposure rose by at least
2 percentage points between 1991 and 2011, as well as all industries ðboth
within and outside of manufacturingÞ for which the predicted full upstream
import exposure measure increased by at least 4 percentage points over
1991–2011.47 Relative to an exposure definition based only on own-industry
import exposure, incorporating upstream linkages expands the exposed
sector to include additional manufacturing industries as well as industries
outside of manufacturing that sell a sizable portion of their outputs toimport-exposed manufacturing firms. For example, the latter group includes forestry, wholesale trade, miscellaneous repair services, and chemicaland fertilizer mining.48 All other industries are designated as nonexposed.
Following our simple example of construction versus chemicals as nonexposed industries, we next subdivide the nonexposed sector into tradables
47
Predicted import exposure is computed from first-stage estimates of eq. ð3Þ
over the single long period 1991–2011.
48
Despite this broad definition of the exposed sector, our regression analysis in
this section will only partially capture the indirect effects working through inputoutput linkages we directly estimated previously. While pairs of industries linked
through input-output relationships tend to co-locate ðe.g., Ellison et al. 2010Þ, many
firms purchase and sell inputs beyond the boundaries of their CZ, and thus any local
strategy will exclude a potentially sizable fraction of these indirect effects.

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S178

Acemoglu et al.

and nontradables. In our nomenclature, tradable industries are those that
produce tradable goods or commodities and specifically constitute the manufacturing, agriculture, forestry, fishing, and mining sectors. We classify allother sectors, including services, as nontradable, though this approach isadmittedly imperfect since some services are also traded.49
Table 7 presents our estimates. The first set of specifications in columns 1–3 pool employment across all sectors to determine the impact ofimport exposure in local labor markets on overall employment. Column 1
considers the relationship between CZ import exposure and changes in
CZ employment-to-population rates without additional controls. The
strongly negative and statistically significant point estimate in this column
indicates that a 1 percentage point increase in the average import penetration of local industries reduces the employment rate among a CZ’s
working-age population by 1.64 percentage points. We refine the estimates and explore robustness in the next pair of columns by controlling
for the initial manufacturing employment share in a local labor market
ðcol. 2Þ and for nine census divisions ðcol. 3Þ. By controlling for local manufacturing intensity, we allow for differential employment trends in the
manufacturing and nonmanufacturing sectors, as we do in our industry-level
estimates of table 6. The controls for census divisions allow for heterogeneity
in regional time trends. Adding these covariates has a modest impact on the
trade coefficient, which remains sizable and statistically significant at 21.70
in column 3.
The regressions of columns 4–6 disaggregate the overall employment
effects of columns 1–3 into their sectoral components. Consistent with theresults of the industry analysis, column 4 shows a strongly negative andstatistically significant effect of import exposure on local labor market
employment in trade-exposed industries. The point estimate indicates thata 1 percentage point increase in local import exposure reduces the shareof a CZ’s working-age population employed in exposed industries by
1.95 percentage points. Between 1999 and 2011, mean CZ import exposure
rose by 1.21 percentage points, while employment in exposed industries declined by 3.64 percentage points of the working-age population. The estimate in column 4 thus implies that 1.32 percentage points ðor 36%Þ ofthis fall can be explained by rising Chinese import competition.50

49
The exposed sector consists of 293 industries ð285 in manufacturing and eightoutside of manufacturingÞ, which together made up 20.2% of 1991 US employment. The nonexposed tradable sector consists of 113 industries ð107 in manufacturing, six outside of manufacturingÞ, making up 6.7% of 1991 employment.
Finally, the nonexposed nontradable sector consists of 73 industries ðall outside
manufacturingÞ accounting for 73.1% of 1991 employment.
50
As above, this calculation discounts the growth of imports by the partial
R-squared of 0.56 of the first-stage regression: 1.32 5 0.56  1.21  1.95.

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Yes
Yes
No
1,444

Yes
No
No
1,444

Yes
Yes
1,444

Yes

21.70**
ð.78Þ

No
No
4,332

.33
ð.39Þ
Yes

2.01
ð.06Þ

21.95***
ð.16Þ

21.95***
ð.62Þ

21.64***
ð.46Þ

Yes
No
4,332

.15
ð.44Þ
Yes

.04
ð.11Þ

22.14***
ð.30Þ

Yes
Yes
4,332

2.01
ð.57Þ
Yes

2.00
ð.11Þ

21.68***
ð.24Þ

Yes
Yes
1,444

Yes

21.89***
ð.65Þ

ð7Þ

Yes
Yes
4,332

2.18
ð.55Þ
Yes

2.05
ð.10Þ

21.66***
ð.19Þ

ð8Þ

Overall Sectoral 1991–2007

NOTE.—Each column reports results from stacking changes in commuting zone employment rates and exposure to Chinese imports over the periods 1991–99 and either 1999–
2011 ðcols. 1–6Þ or 1999–2007 ð7–8Þ. In cols. 1, 2, 3, and 7, the dependent variable is 100  the annual change in the ratio of total employment to working-age population ðN 5 1,444 5
722 commuting zones  two periodsÞ. In the other columns, the dependent variable is 100  the annual change in the ratio of sectoral employment to working-age population, withindustries partitioned into three sectors: industries exposed to trade competition, nonexposed industries that produce tradable goods, and all remaining nonexposed industries ðN 54,332 5 722 commuting zones  three sectors  two periodsÞ. See the text for details. Commuting zone import exposure is an employment-weighted average of annualized changes in
exposure to Chinese imports within local industries; it is instrumented as described in the text. Employment is computed in the CBP; population data come from the Census
Population Estimates. The manufacturing share of baseline commuting zone employment is computed in 1991 ðfor the 1991–99 periodÞ or 1999 ðfor the 1999–2011 and 1999–2007
periodsÞ. Census division dummies control for nine census divisions. Observations are weighted by 1991 commuting zone population. Standard errors in parentheses are clustered oncommuting zone.
** p < .05.
*** p < .01.

Sector  time effects
Sector  manufacturing employment
share at baseline
Sector  census division dummies
Observations

Commuting zone import exposure 
1fnonexposed nontradable sectorg

Commuting zone import exposure 
1fnonexposed tradable sectorg

Commuting zone import exposure 
1fexposed sectorg

Commuting zone import exposure

ð6Þ

ð4Þ

ð5Þ

Sectoral Employment 1991–2011

ð2Þ

ð1Þ

ð3Þ

Overall Employment 1991–2011

Table 7
2SLS Estimates of Import Effects on Commuting Zone Employment-to-Population Ratios

S180

Acemoglu et al.

As our conceptual discussion anticipates, the estimate in column 4 also
shows some offsetting employment growth in nonexposed industries,
corresponding to the net impact of local reallocation and Keynesian demand effects. However, the offsetting employment effect is substantiallysmaller than the employment reduction in exposed industries and is neverstatistically significant. These estimates suggest that employment gains
through the sectoral reallocation effect are largely offset by negative aggregate demand effects. In parallel with our specifications examining overall employment impacts, we refine the estimates in the next pair of columns by controlling for initial local labor market manufacturing intensity
ðcol. 5Þ and census divisions ðcol. 6Þ, with the coefficients on these controls
allowed to vary by sector. Adding these covariates only modestly changes
the estimated negative impact of import exposure on employment in exposed industries, while the small and imprecise estimates for offsetting
employment gains decline to almost zero. The final columns replicate thespecifications from columns 3 and 6 over the stacked periods 1991–99 and
1999–2007. The results are similar to those for the full sample period andsuggest negative effects of trade competition on employment in exposedindustries, combined with small and insignificant effects in nonexposed
sectors.
While our estimates suggest the presence of strong aggregate demand
effects that limit employment gains in the nonexposed sectors of tradeexposed local labor markets, we would anticipate that these local demand
effects primarily have an impact on employment in the nontraded sector
rather than the nonexposed tradable sector. Our results, however, provide
scant evidence for differential employment impacts in the two nonexposed sectors. In columns 4 and 5, the point estimates for nontradables
exceed the point estimate for nonexposed tradables; in columns 6 and 8,
the relationship is reversed.
Why does reallocation fail to accord more clearly with the simple reasoning outlined in Section VI.A? It is conceivable that the small increase inemployment in nontradable sectors detected in columns 4 and 5 ðthough
not in col. 6Þ may be related to the rapid rise in the US aggregate trade deficit
during our sample period, a substantial part of which reflects a growing
trade imbalance with China ðfig. 2Þ. In response to import competition, an
open economy normally reallocates resources out of some tradable industries into others, at least under balanced trade. If, however, the trade
shock is accompanied by a rise in the trade deficit, then the reallocationfrom exposed tradables into nonexposed tradables may be delayed, shifting
employment into nontradables instead; that is, the deficit may fuel increasing expenditure in the domestic economy, part of which falls on
nontradable consumption. While this reasoning is not inconsistent with along-run reallocation toward nonexposed tradables, the large and growing

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Import Competition and the Great US Employment Sag

S181

US trade deficit during the period under study may have significantly
slowed down such a reallocation. This reasoning is, unfortunately, silent
on why a rising US trade deficit coincided with China’s growing importpenetration. It nevertheless underscores that shifts in global imbalances
may complicate the simple adjustment mechanism we posit.
Quantitatively, the estimates in column 6 of table 7 encompass four
impacts of Chinese trade competition on local labor market employment:
direct employment effects in exposed industries, indirect employment
effects via local input-output linkages between industries, local reallocation effects, and local aggregate demand effects. As summarized in table 8,
the coefficient estimates imply that had import competition from China
not increased after 1999, trade-exposed industries in local labor markets
would have avoided the loss of 2.35 million jobs. Comparing this quantity
to the outcome of our national industry analysis, it is modestly larger thanthe employment effect derived from panel B of table 6 reported above,
which incorporated both the direct and the upstream effects of importcompetition and tallied employment reductions in trade-exposed manufacturing and nonmanufacturing industries at 1.98 million jobs. The factthat employment effects on exposed industries in CZs are slightly larger
than the direct and indirect effects of import competition in national
industries is suggestive of negative local aggregate demand spillovers. Such
spillovers imply that multipliers operating at the local level suppress
demand in nonexposed industries as well, inducing further employment
declines in trade-exposed industries.
Our estimates imply near zero, though imprecisely estimated, employment effects of trade exposure on nonexposed industries. In the absence offurther increases in import penetration from China after 1999, the resultssummarized in table 8 show that nonexposed industries would have shed
18,000 fewer jobs. Combining figures from exposed and nonexposed industries, the overall local impact is 2.37 million jobs whose loss would havebeen averted without further increases in Chinese import competition after
1999. With the numerous caveats acknowledged, our conceptual framework in Section II suggests that this estimate is a lower bound on the aggregate total impact of increased import competition from China on national
employment. In particular, this estimate does not include the components
of industry interlinkage effects and aggregate demand effects that work at
the national level. This lower-bound estimate is relatively close to the jobs
lost on the basis of our industry-level analysis in panel B of table 6 ðshownin table 8Þ, which combines direct competition effects and interindustry
linkages with nonmanufacturing sectors. Recall that table 6’s industry-level
estimate of the jobs lost does not include reallocation and aggregate demand
effects. Since our analysis in this section indicates that employment losses
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Industry
Industry

Industry

Commuting zone

Table 2, cols. 3/4
Table 6, panel A,
cols. 6/9

Table 6, panel B,
cols. 6/9

Table 7, cols. 6/9

Effect of local import exposure on
employment in the commuting zone,
controlling for baseline manufacturing
share and for census divisions

Direct and “full” ðhigher-orderÞ upstream
effects of import exposure

Direct effect of import exposure
Direct and “first-order” upstream
effects of import exposure

Description
Manufacturing
Total
Manufacturing
Nonmanufacturing
Total
Manufacturing
Nonmanufacturing
Total
Exposed industries
Nonexposed tradables
Other nonexposed

Affected SectorðsÞ

1999–2011
2560
21,5812928
2653
21,9792985
2994
22,36722,348
21
217

2277
2556
2404
2152
2645
2421
2224
2743
2737
0
25

2837
22,137
21,3322805
22,624
21,406
21,218
23,11023,08621
223

1991–2011

2853
22,21821,4142804
22,66921,47521,19423,031
22,663279
2289

1991–2007

Implied Employment Changes ð000sÞ
1991–99

NOTE.—Reported quantities represent the change in employment attributed to instrumented changes in import exposure in each of our preferred specifications. Negative values
indicate that import exposure is estimated to have reduced employment. For the industry-level analyses, we first use the estimated coefficients to predict the changes in eachindustry’s log employment induced by changes in import exposure over the periods 1991–99 and 1999–2011. Concretely, we multiply the coefficient of interest by the observedchange in import exposure, then multiply this product by .56 ðthe partial R-squared from our baseline first-stage regressionÞ. We then use each industry’s observed end-of-period
employment to convert these estimates from logs into levels. Upstream effects are handled similarly. For the commuting zone analyses, we first use observed changes in imports per
worker—again discounted by .56—to predict the trade-induced change in each commuting zone’s employment-to-population ratio within the indicated sectors over the periods
1991–99 and 1999–2011. We then multiply by end-of-period commuting zone working-age population to compute the implied changes in each sector’s employment in each
commuting zone. Summing these sectoral estimates across commuting zones yields nationwide estimates. See the text for definitions of the exposed, nonexposed tradable, andnonexposed nontradable sectors. For both industry-level and commuting zone–level analyses, predictions for 1991–2011 equal the sum of the predictions for the two subperiods.
Predicted employment changes for the period 1991–2007 are computed similarly, using coefficients from models estimated over the stacked periods 1991–99 and 1999–2007.

Unit of Analysis

Specification

Table 8Implied Employment Changes Induced by Changes in Exposure to Chinese Imports

Import Competition and the Great US Employment Sag

S183

to reallocation effects, our industry-level estimates of employment reduction should indeed be lower bounds.51
VII. Concluding Discussion
In the years leading up to the Great Recession, overall US employment
growth was slow and manufacturing employment experienced a steep
contraction. In this article, we investigate the contribution of the rise inimport competition from China to this employment “sag.”We begin by estimating the direct effect of trade competition on employment in manufacturing industries that are differentially exposed to
growing Chinese import penetration and then expand the analysis to include multiple general equilibrium channels through which trade exposure
may affect employment: other sectors might be affected because they are
related to the affected sectors through input-output linkages; employment
may reallocate away from trade-exposed industries toward nonexposed industries; and Keynesian-type aggregate demand spillovers may significantly
magnify the direct competition effect.
In our analysis of US national industries, we estimate upstream anddownstream trade effects for both manufacturing and nonmanufacturing sectors. We expect upstream effects to contribute to further job losses,
while the impact of downstream effects is ambiguous. Consistent with these
expectations, we find large negative employment responses when an industry’s customers are exposed to trade competition and unstable effects
when an industry’s suppliers are exposed to trade competition.
As a complementary strategy, we assess the impact of Chinese trade on
US commuting zones to jointly estimate reallocation and aggregate demand
effects at the local level. Theoretically, if an industry contracts in a local labormarket because of Chinese competition, then, barring substantial interregional migration, some other industry in the same labor market should expand. In addition, part of any aggregate demand spillovers will also accrue
to the local labor market. Our estimates show sizable job losses in exposedindustries and few, if any, offsetting job gains in nonexposed industries, apattern that is consistent with substantial job loss due to aggregate demand
spillovers.
Our results are a first step in quantifying the employment impact ofincreasing import competition on the US labor market. Several questions
remain unanswered that could be addressed in future work. Using plantlevel data to achieve a finer distinction between tradable and nontradable
51

In particular, recall that the industry-level numbers could underestimate the net
employment losses due to aggregate demand effects or overestimate these losses due
to reallocation effects. But if reallocation effects are modest and are swamped bydemand effects at the local level, as suggested by the table 7 estimates, we would also
expect the demand effects to dominate at the aggregate level—especially since thesedemand effects are themselves underestimated at the local level.

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S184

Acemoglu et al.

industries would enable both a sharper test of the implications of local
general equilibrium interactions and a separate quantification of reallocation and aggregate demand effects. We should in particular see employment declines in nontradables due to local aggregate demand spillovers, but no differential decline in tradables except through geographicallyconcentrated input-output linkages. This perspective could elucidate how
local and national labor markets respond to growing import competition,in particular, allowing us to determine to which degree shocks propagate
locally or at the national level.
We finally note that, though our article has focused on the contributionof rising international competition to the US employment sag of the
2000s, we have had comparatively less to say about the impact of tradeduring the Great Recession. As shown in figure 2, US imports from China
dropped sharply in 2009. This might imply that exporters to the UnitedStates—China in particular—absorbed part of the demand shock accompanying the Great Recession that would otherwise have further reduced
US employment ðalbeit from a notionally higher baseÞ. While this hypothesis is intuitive, additional exploration of US manufacturing data suggests
otherwise. We find that US manufacturing industries that were heavily
exposed to Chinese import competition during the 1999–2007 period continued to see rapid, differential employment declines during 2007–11, despite the fact that there was almost no correlation between industry-levelchanges in trade exposure during 1999–2007 and changes in trade exposure
during 2007–11.52 This pattern suggests that the trade shocks of the priordecade cast a long shadow over US manufacturing, even when trade pressure eased temporarily. One explanation for this long shadow is that US
manufacturers recognized that the loss in comparative advantage in the
sectors that China had penetrated in the prior decade was largely permanent
whereas the lull in trading activity was temporary. Indeed, as shown infigure 2, US imports from China more than made up all of their ground lostin 2009 by the following year and then rose further from there. Thus, tradepressure appears to have contributed to the US employment sag not justbefore but also during the Great Recession, despite the temporary drop-off
of international trading activity during this period. Although much evidence
When we regress 100  the annual log change in manufacturing industry
employment between 2007 and 2011 on changes in Chinese import competition
between 2007 and 2011 and between 1999 and 2007 ðexpressed as percentage
points of 1991 US market volumeÞ, we find52

c j;07–11 5 25:022 1:06  DIPj;99–07 1 0:59  DIPj;07–11 :DL
ð0:52Þ

ð0:40Þ

ð0:67Þ

This substantial impact of Chinese import competition between 1999 and 2007 on
2007–11 employment growth suggests a pattern of delayed declines in employment in affected industries. We obtain similar results if we control for 10 one-digit
sector dummies.

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Import Competition and the Great US Employment Sag

S185suggests that rising labor costs in China augur a reduction in trade pressurein the years ahead ðLi et al. 2012Þ, our analysis suggests that this particular
Chinese export has yet to reach US shores.
Appendix A
Additional Results
Table A1
Industry-Level Control Variables
Production workers’ share of employment, 1991
Ratio of capital to value added, 1991
Log real wage ð2007 US$Þ, 1991
Computer investment as share of total, 1990
High-tech equipment as share of total investment, 1990
Change in industry share of total employment, 1976–91
Change in log real wage, 1976–91

Mean

SD

Min

Max

68.43
.92
10.54
6.56
8.24
2.03
3.57

15.50
.55
.29
6.07
4.84
.07
9.94

18.72
.19
9.78
.00
1.20
2.42
232.01

97.62
3.52
11.09
43.48
18.25
.07
48.06

NOTE.—N 5 392 four-digit manufacturing industries. Observations are weighted by industry employment in 1991, as measured in the CBP. Production workers’ share, the ratio of capital to value added, log real
wage, and the changes in industry employment share and in log real wage are computed using the NBERCES Manufacturing Industry Database; total employment in 1976 and 1991 is computed from the Current
Employment Statistics. The remaining control variables are taken from Autor et al. ð2014Þ. Share variables
are expressed in percentage points.

Table A2
Estimates of Import Effects on Log Gross Output and Log Price Deﬂators
NominalShipments
ð1Þ

Real
Shipments
ð2Þ

Shipments
Deflator
ð3Þ

A. All Manufacturing Industries ðN 5 768Þ
100  annual D in US exposure
to Chinese imports
One-digit manufacturing sector controls

21.08***
ð.32Þ
Yes

2.17
ð.44Þ
Yes

2.91**
ð.42Þ
Yes

B. Exclude Computer Industries ðN 5 712Þ
100  annual D in US exposure
to Chinese imports
One-digit manufacturing sector controls

21.00**
ð.47Þ
Yes

2.86**
ð.41Þ
Yes

2.14*
ð.08Þ
Yes

NOTE.—Each column stacks changes in the indicated outcome and changes in US exposure to Chinese
imports over the periods 1991–99 and 1999–2009. In panel A, the sample consists of 384 four-digit
manufacturing industries for which data are consistently available in the NBER-CES Manufacturing
Industry Database ðN 5 768 5 384 industries  2 periodsÞ. In panel B, we exclude 28 computer-producing
industries corresponding to NAICS 334 ðN 5 712 5 356 industries  2 periodsÞ. The dependent variablein each column is 100  the annual log change in the indicated outcome, as computed in the NBER-CES.
The change in US exposure to Chinese imports is instrumented as described in the text. All specifications
include time effects as well as controls for 10 one-digit manufacturing sectors. Observations are weighted
by 1991 employment in the NBER-CES. Standard errors in parentheses are clustered on three-digit
industries.
* p < .10.
** p < .05.
*** p < .01.

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Table A3
Direct, Upstream, and Downstream Import Exposure, 1991–2011
Manufacturing Industries
ðN 5 392Þ
Mean/
SD Median Min
Direct import exposure:
Direct exposure
Instrument for direct
exposureFirst-order indirect
exposure:
Upstream exposure
Instrument for upstream
exposure
Downstream exposure
Instrument for
downstream exposure
Full ðhigher-orderÞ indirect
exposure:
Upstream exposure
Instrument for upstream
exposure
Downstream exposure
Instrument for downstream
exposure

Max

Nonmanufacturing
Industries ðN 5 87Þ
Mean/
SD Median Min Max

.50
ð.94Þ

.14

2.02 10.93

.44
ð.76Þ

.15

2.52

8.59

.16
ð.26Þ

.06

.00

1.88

.03
ð.04Þ

.01

.00

.19

.12
ð.18Þ
.10
ð.11Þ

.05

.00

1.55

.01

.00

.22

.07

.00

.83

.02
ð.03Þ
.03
ð.04Þ

.02

.00

.24

.09
ð.08Þ

.07

2.02

.46

.02
ð.03Þ

.02

.00

.14

.24
ð.35Þ

.09

.00

1.98

.06
ð.07Þ

.03

.00

.67

.19
ð.25Þ
.14
ð.13Þ

.10

.00

1.61

.02

.00

.65

.11

.00

1.05

.05
ð.06Þ
.05
ð.05Þ

.04

.01

.33

.14
ð.10Þ

.12

2.01

.61

.05
ð.04Þ

.04

.01

.21

NOTE.—The direct import shock to industry i is defined as 100  the annual change in US exposure to
Chinese imports in that industry over 1991–2011. The first-order measure of upstream ðrespectively,
downstreamÞ import exposure experienced by i is a weighted average of the direct import exposure
experienced by its customers ðsuppliersÞ j, where the weight on industry j equals i’s sales to ði’s purchasesfromÞ j divided by i’s total sales. The full upstream and downstream exposure measures are constructedusing the Leontief inverse of the input-output matrix to incorporate higher-order linkages; see the text for
details. Instruments for the direct, upstream, and downstream exposure measures are constructed analogously, using changes in comparison countries’ exposure to Chinese imports in own and linked
industries. Observations are weighted by 1991 industry employment in the CBP.

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Table A4
Changes in Commuting Zone Import Exposure and Employment-to-Population
Ratios1991–99Mean/
SD Median
D in local exposure to Chinese
imports:
100  annual D in commuting zone exposure
to Chinese imports
Instrument for D in
commuting zone exposure
to Chinese imports
D in employment/workingage population:
100  annual D in overall
employment/population

1999–2011

Min

Max

Mean/
SD Median

Min

Max

.05
ð.05Þ

.04

.00

.95

.10
ð.07Þ

.09

.00

.69

.04
ð.04Þ

.04

2.06

.53

.13
ð.09Þ

.12

2.01

.79

.73
ð.39Þ

.73

21.15 3.48 2.52
ð.40Þ

2.58

22.16 2.63

2.04

21.90 1.21 2.30
ð.17Þ

2.30

21.55

2.04

2.70 1.47 2.07
ð.08Þ

2.08

2.85 1.52.82

2.62 3.21 2.14
ð.32Þ

2.14

21.82 1.34

100  annual D in employment/population within
exposed industries
2.03
ð.16Þ
100  annual D in employment/population within
nonexposed tradable
industries
2.04
ð.10Þ
100  annual D in employment/population within
other nonexposedindustries
.80
ð.32Þ

.64

NOTE.—N 5 722 commuting zones. The annual change in commuting zone exposure to Chinese imports
is a weighted average of changes in US import exposure in 392 four-digit manufacturing industries, where
the weights are start-of-period employment shares within the commuting zone. The instrument is constructed by replacing US imports from China with imports from China by a set of comparison countries andby using 1988 commuting zone employment shares as weights; see the text for details. Imports are deflated toconstant dollars using the PCE price index. In the second panel, each variable describes the annual change
in 100  total or sectoral employment divided by the commuting zone population between the ages of 15 and64. Exposed industries include manufacturing industries for which the predicted increase in Chinese importpenetration exceeds 2 percentage points between 1991 and 2011, plus industries for which the predicted
increase in the measure of full upstream import exposure ðincorporating higher-order linkagesÞ exceeds
4 percentage points over 1991–2011. Among nonexposed industries, we define agriculture, forestry, fishing,mining, and manufacturing industries as tradable and all other industries as nontradable. Employment iscomputed in the CBP, and population is computed using the Census Population Estimates. Observations areweighted by total 1991 commuting zone population.

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Acemoglu et al.

FIG. A1.—First-stage regression, 1991–2011. Each point represents a four-digit
manufacturing industry ðN 5 392Þ. The change in US exposure to Chinese imports
is defined as the change in US imports from China divided by 1991 US market
volume; the change in the comparison countries’ exposure to Chinese imports isdefined as the change in these countries’ imports from China divided by 1988 US
market volume. Lines are fitted by OLS regression, weighting by each industry’s
1991 employment in the CBP. The 95% confidence interval is based on standard
errors clustered on 135 three-digit industries. The slope coefficient is .98 withstandard error .14; the regression has an R-squared of .62. A color version of thisfigure is available online.

Appendix B
Derivation of the Downstream and Upstream Effects
In this appendix, we briefly outline the justification for the specifications we use for the upstream and downstream effects in Section V of thearticle.
Setup
Consider a static perfectly competitive economy with n industries, andsuppose that each industry j 5 1, . . . , n has a Cobb-Douglas production
function of the formalyj 5 lj jPna

xjiji :

i51

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ðB1Þ

Import Competition and the Great US Employment Sag

S189

Here xji is the quantity of goods produced by industry i used as inputs byindustry j. We assume that, for each j, alj > 0, and aji 0 for all i, and thatnalj 1 o aji 5 1;
i51

so that the production function of each industry exhibits constant returns
to scale. ðPhysical capital can also be introduced without affecting theresults, but we omit it to simplify the notation and the discussion.Þ
The output of each industry is used as input for other industries or
consumed in the final-good sector. In addition, there are also imports
from abroad ðsay ChinaÞ, and we ignore exports for simplicity ðand thus
also ignored is the trade balance conditionÞ. The market-clearing condition for industry j can then be written asnyj 5 cj 1 o xkj 2 mj ;

ðB2Þ

k51

where cj is final consumption of the output of industry j, and mj denotes
total ðrealÞ imports.
The preference side of this economy is summarized by a representative
household with a utility function
uðc1 ; c2 ; : : : ; cn Þ:
We focus on the competitive equilibrium of this economy.
Main Result
First consider the unit cost function of sector j:PnlCðp; wÞ 5 Bj wajapi ji ;

ðB3Þi51

where p is the vector of prices, w is the wage rate, and

Bj 5

1
alj

P

alj n 
i51

1
aji

aji

is a sector-specific constant.
Cost minimization of industry j ðgiven competitive marketsÞ implies thataji 5

pi xji
;
pj yj

ðB4Þ

where pj is the price of the output of industry j. This expression makes it
clear that aij’s also correspond to the entries of the input-output matrix,
which we denote by A.

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Next, given the constant returns to scale production function of eachsector specified in ðB1Þ, prices satisfy the zero profit conditions of the nsectors in the competitive equilibrium. In particular, the price of good jmust be equal to the unit cost function of that sector, ðB3Þ, and thusPnlpj 5 Bj wajapi ji :

i51

Taking logs, we havenln pj 5 ln Bj 1 alj ln w 1 o aji ln pi

for all j ∈ f1; : : : ; ng:i51

Let us choose w 5 1 as the numeraire. Then, these equations define ann-equation system in n prices
ln p 5 ðI 2 AÞ21 b;
where, as noted above, A is the input-output matrix of the economy, and b
is the vector with entries given by ln Bi ðand we are using the fact that ln
w 5 0Þ. This implies that prices in this economy are determined independently of imports ðpurely from the supply sideÞ. Consequently, there
will be only quantity responses to imports.
But from consumer maximization, with unchanged prices, relative
consumption levels remain unchanged. How total consumption is affected
depends on whether there is trade balance or not. With trade balance, theeconomy would have to export some goods to make up for the increasein imports. Here for simplicity, we allow for a trade deficit and thus leave
the entire consumption vector unchanged. With unchanged consumption
levels, we must have from ðB2Þ combined with ðB4Þ thataji dð pj yj Þ 5 dð pi xji Þ:ðB5ÞFor future reference, let us define nominal values ðwhich are more
useful for several of the expressions belowÞ with tildes. For example,
x~ji ; pi xji ;y~j ; pj yj ;~ j ; p j mj :mThen ðB5Þ can be equivalently written as
yj 5 d~xji :aji d~

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ðB6Þ

Import Competition and the Great US Employment Sag

S191

Now totally differentiating the resource constraint, ðB2Þ, for sector i,we obtainndyi 5 dci 1 o dxji 2 dmi ;j51which, using ðB6Þ and the fact that consumption levels are not changing,can be written asndð pj yj Þdð pi yi Þdð pi mi Þ
5 o aji
2
:
pi yi
pi yi
pi yi
j51

Writing this in matrix form, and noting that, because prices are constant,dð pi yi Þ=pi yi 5 d ln yi , we have^0 d lny 2 Ldm^
d ln y 5 A
0 21^^
5 2ðI 2 A Þ Ldm

ðB7Þ

^ Ldm;
^
5 2H
0





^ 21 ,^ 5 ðI 2 AÞ
^ is the vector with entries given by pimi, H
where m01
a^11 a^12
B a^21 a^22
C
B
C
^ 5B
C;
A
B
C
@
A
a^nn
with entries a^ij 5 pj xij =pj yj ðas opposed to aij, which is equal to pjxij/piyiÞ,and L is the matrix with 1=~
yj on the diagonals and zero on the nondiagonals.53 Intuitively, any import shock creates a direct negative effecton the directly affected sector, which is captured by the matrix L, and the^ 0.
indirect effects are summarized by the Leontief inverse matrix H
We can see from this expression that there will be only upstream effects^ A^ 0 , that matters in theðsimply note that it is the transpose of the matrix A,
Leontief inverse, thus corresponding to transmission only in the upstream
directionÞ. This is a consequence of the fact that there are no changes inprices, and hence quantities will respond to changes in imports; but foreach change in the quantity of a sector directly affected by imports from
China, the quantities of inputs that it receives from its suppliers will have
^ 0 is exactly
to adjust, causing upstream propagation.54 In fact, the matrix H^ is less than one, I 2 A^ 0 is invertible.
Note that since the largest eigenvalue of A
There would be further effects if we were to impose trade balance, becausesome sectors would have to expand in order to compensate for the increase inimports. In that case the matrix L would have nonzero off-diagonals.
53
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Acemoglu et al.

what we use in Section V for computing the full ðLeontief inverseÞ upstream effects.
Equation ðB7Þ gives the output responses to import shocks. It is
straightforward to derive from this the employment responses, which are
our main focus. In particular, given the Cobb-Douglas form of the production function in ðB1Þ, cost minimization for industry i implies thatwli 5 ali pi yi . Since the wage is constant, employment in industry i is proportional to its nominal output, enabling us to work with an analogue ofðB7Þ with employment on the left-hand side.
We next develop a more heuristic derivation of this result, which provides further intuition, shows how the full effects summarized by the^ 0 come about, and also explains why under more
Leontief inverse matrix H
general conditions there might also be some downstream effects.
Heuristic Derivation
Let us first ignore the second- and higher-order input-output linkages
and focus on first-order impacts. Let us use the notation for nominal
variables introduced above and begin by approximating the impact of theincrease in imports in industry j on domestic production in the same
~ j . ðThis is clearly an approximation, since as ourindustry as d~yj ≈ 2dm
derivation in the previous section showed, there will be higher-order effects^ 0 .Þ
on the output of sector j as captured by the Leontief inverse matrix H
Note further that from ðB4Þ, any reduction in the value of output of anindustry translates into a proportionate reduction in all of the inputs, inparticular,
d~
xji
5 aji
d~
yj

ðB8Þ

for each industry i. Then from ðB8Þ we have
d~
yi
d~
yi
≈ 2
5 2aji~jdmd~yj
for each industry i ≠ j, and we haved~yj
≈ 2ð1 1 ajj Þ~jdm
for industry j itself, reflecting both direct import substitution and theresultant decline in j’s demand for its own inputs. These two cases can be
dealt with succinctly by defining dij ; 1fi 5 jg, so that for any industries iand j,

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Import Competition and the Great US Employment Sag

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d~
yi
≈ 2ðdij 1 aji Þ:~jdm
For small changes in mj, a first-order Taylor approximation gives thetotal impact on domestic production in industry i as
d~
yi ≈

d~
yi
 dm
~ j ≈ 2ðdij 1 aji Þ  dm
~ j:
~j
dm

Now turning this into a proportional ðlogÞ effect by normalizing theimpact on industry i relative to its domestic production, we obtain
d~
yi
d~
yi
1
1
 dm
~ j  ≈ 2ðdij 1 aji Þ  dm~j  :
≈
~j
dm
y~i
y~i
y~i
This expression shows how industry i is affected when a single industry
j to which it sells inputs is exposed to import competition. We can nextcompute the total effect on industry i from the full vector of import changes
by summing this expression across all of i’s customer industries:

 n
d~
yi
1
 dm
~j 
ðd ln~
yi Þfirst-order ≈ o~jy~i first-orderj51 d mn~j 
≈ 2 o ðdij 1 aji Þ  dm
j51
n

~j 
5 2 o ðdij 1 a^ji Þ  dm
j51

1
y~i

ðB9Þ

1
;
y~j

^ used in equation ðB7Þ.
where a^ij ’s correspond to the entries of the matrix A
Now using the same matrix notation as in that equation, this relationshipcan be written as^ 0 ÞLDm;^
dln yfirst-order ≈ 2ðI 1 A
which clarifies that first-order effects take exactly the same form as the full
effects we just derived, but with only the direct effect working through the^ included ðhence the first-order designationtranspose of the matrix A
rather than the full effectsÞ. This expression is what we use to computefirst-order downstream effects in Section V.55
55

~ j  1=~
Using ðB4Þ, we can rewrite oj51 a^ji  dm
yj as
n

~j
x~ji dm
;
~
y
i
j
j51
n

o y~

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Acemoglu et al.

Our more rigorous derivation in the previous subsection makes it clearthat the first-order effect cannot be isolated from higher-order effects,
since an increase in mj will have an impact on yk and from there on the
sectors supplying inputs to k and so on. Letting Ai denote the ith columnof A, A2i denote the ith column of A2, and so on, we can obtain
 n


d~
yi
1
 dm
~j 
ðd ln~
yi Þfull 5 o~jy~i full
j51 d m


1
1
1
0
0
0
~  1 ðAi Þ  dm
~  1 ðA2i Þ  dm
~  1 5 2 ei  dm
y~i
y~i
y~i
0

0

0

~ 
5 2½ei 1 ðAi Þ 1 ðA2i Þ 1     dm0

~ 
5 2½ðI 2 A Þ21i   dm

1
y~i

1
:
y~i

Using the same notation as above, this can be rewritten as^ 0 Þ21 Ldm^
d lny 5 2ðI 2 A
0^ Ldm;^
5 2H
confirming ðB7Þ.
Downstream Effects
Downstream effects simply correspond to effects that spread downstream following the input-output matrix A, and in our empirical work
^ 0 Þ21 Ldm^
we construct first-order and full downstream effects as 2ðI 2 A
21^ respectively.
and 2ðI 2 AÞ Ldm,
The above derivation confirms that, in our baseline model, there are nodownstream effects from changes in imports. This result, however, depends
on certain assumptions. First, the focus on competitive equilibrium inwhich there are no relationship-specific investments between input suppliers and customers is important. Second, the feature that there are no priceeffects, which will no longer be true with departures from perfectly competitive markets, also plays a major role.
which clarifies that the upstream effect on industry i is a sales-weighted average ofthe proportional import shocks experienced by its customers j. In our empiricalwork, import changes correspond to changes in Chinese import penetration, andthe weights are constructed using the 1992 BEA benchmark input-output table.
Our empirical measure also denominates import changes by US market volume ineach industry ðshipments plus imports minus exportsÞ rather than by industry
shipments.

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In particular, without the competitive equilibrium assumption, the increase in imports may drive some producers out of the market, and this may
have a negative impact on firms that are their customers, creating negative
downstream effects. Conversely, if there are declines in the prices of goods
being imported more intensively from China, this may create positive
downstream effects as customers using these goods as inputs can expand
their operations.
Ultimately, whether there are downstream effects or not is an empiricalquestion, and our results do not provide much evidence for sizable downstream effects.
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