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This is the paradox of sin taxes, the class of taxes that includes tobacco. These extra dollars and cents levied on products and activities considered detrimental to consumers—traditionally tobacco, alcohol, and gambling—are intended to accomplish two contradictory goals: Like all taxes, they generate revenue for the taxing entity, but they also aim to deter the behavior being taxed—which can ultimately negate the first goal.

Indeed, research suggests that higher cigarette prices play a role in curbing smoking.4 A reduction in smoking is good for public health, but it has been so significant that inflation-adjusted tobacco tax revenue nationwide has waned in recent years despite frequent tax rate hikes. Revenue from tobacco taxes depends on high rates of smoking—which the taxes discourage—making these taxes an unreliable source of long-term revenue, especially for funding recurring budget items.

This study concludes unambiguously that the tax imposed on gambling activities in Switzerland is regressive. Lower income individuals contribute proportionally more to the total state revenues than do higher income groups. Indeed, the Suits index shows a negative value of –0.184, indicating a clear regressive tax pattern. The regression analysis corroborates this first result. The income elasticity of 0.199 is clearly and statistically lower than one, providing further proof of a regressive tax pattern. Decreasing income by 10% results in a decrease of only 1.99% in the gambling budget. Therefore, the economic incidence clearly shows that the final distribution of the tax burden is proportionally larger for individuals with lower incomes. Thus, this tax violates one of the main qualities of a good tax in Switzerland, namely vertical equity. Vertical equity suggests that individuals with higher income should have a larger tax burden. However, this is obviously not the case in the present study.

Another type of inequality is related to educational level. In the regression analysis, this coefficient is found to be negative. This pattern shows that the more educated an individual is, the less money he/she will spend on gambling. According to our analysis, state gambling revenues are expected to rise if the population that has access to gambling is poorly educated. These two results lead to the surprising conclusion that the revenues to the state from games of chance will be higher if the population is poor and not well educated. This finding is in line with the findings of Abt, Smith, and Christiansen (1985).

The second reason is that in many states, when cigarette tax rates are increased sizably above neighboring states, consumers are more inclined to buy cigarettes elsewhere. Cigarette smugglers are known to buy cigarettes in low-tax jurisdictions to resell them illegally in high-tax jurisdictions, maintaining a competitive advantage over the legal market by not collecting the excise tax. Cigarette smuggling is especially prominent in New York, where the tax is $4.35 per pack (with an add-on tax in New York City). Approximately  of cigarettes sold in New York are smuggled in from another location, with states like California, Washington, New Mexico, and Minnesota falling not far behind. This cross-border effect considerably reduces the productivity (from a revenue standpoint) of cigarette taxes over time.

Because cigarette tax revenue is so unstable, policymakers and taxpayers should be skeptical of proposals to tie cigarette tax increases to the funding of , as collections are likely to decline over time.

Project 21, from the National Leadership Network of Black Conservatives, issued the following news release:

Excise taxes should be repealed because they harm the poor disproportionately by requiring them to pay a higher portion of their incomes in taxes, say members of the Project 21 black leadership network. In its "Blueprint for a Better Deal for Black America," Project 21 calls for repealing both gasoline taxes and "sin taxes" on items such as alcohol, soda, tobacco, non-tobacco nicotine products and fatty foods. This, they say, will reduce the burden on those who are economically at risk.

Sin taxes (narrow taxes on specific goods or industries) are often justified as a key legislative tool to reduce unhealthy behaviors such as the use of tobacco and excessive alcohol consumption (Cawley & Ruhm, 2011). Research suggests these taxes are generally effective in decreasing the purchase and consumption of targeted products (Chaloupka & Warner, 2000; Xu & Chaloupka, 2011). In addition to directly impacting demand, however, sin taxes have other spillover effects. For example, Cook and Moore (1993) observe that beer-excise tax rates are negatively correlated with rape and robbery at the state level. Beatty et al. (2009) document that large tax differentials near borders induce cross-border shopping and smuggling of alcohol and tobacco. Rozema and Ziebarth (2017) find that smokers respond to increases in cigarette taxes by increasing their use of food stamps. Our study adds to the literature on the spillover effects of sin taxes by examining whether the taxpayers targeted by the tax are more likely to defraud other individuals in unrelated transactions. In general, this adds to our understanding of the true effects of sin taxes, and studying the forces that lead to fraud are particularly important because fraud can erode trust in markets.

There are two reasons why affected taxpayers might commit more fraud following a rise in a sin tax. First, tax increases can reduce the buying power of targeted taxpayers. Given the importance of incentives in spurring fraud (e.g., Charness et al., 2019; Dimmock & Gerken, 2012; Kajackaite & Gneezy, 2017), this reduction in purchasing power could increase an affected taxpayer's marginal value of income and, accordingly, their associated incentive to commit fraud. Second, being treated unfairly makes cheating behavior easier to rationalize, even when the potential victims of fraud are not the party behind an individual's unfair treatment (Houser et al., 2012). Surveys consistently report that affected taxpayers view sin tax hikes as unfair (Carlson, 2005; Dugan, 2014; Saad, 2002), which suggests that these affected taxpayers may find it easier to rationalize defrauding others in the aftermath of sin tax hikes.

WASHINGTON – So-called sin taxes are a useful tool for supporting public health objectives and are sometimes used for budget revenues, but their sustainability as a revenue source is a mixed picture, according to a new study by Pew Charitable Trusts and the Rockefeller Institute of Government.

States increasingly are looking to sin taxes as short-term budget fixes by enacting new taxes on e-cigarettes and marijuana in jurisdictions where it’s legal, said Mary Murphy, the project director at Pew. Sin taxes also may be used to back municipal bonds.

Because lower-income households spend a greater share of their income on consumption than higher-income households do, the burden of a VAT is regressive when measured as a share of current income: the tax burden as a share of income is highest for low-income households and falls sharply as household income rises. Because income saved today is generally spent in the future, the burden of a VAT is more proportional to income when measured as a share of income over a lifetime. Even by a lifetime income measure, however, the burden of the VAT as a share of income is lower for high-income households than for other households. A VAT (like any consumption tax) does not tax the returns (such as dividends and capital gains) from new capital investment, and income from capital makes up a larger portion of the total income of high-income households.

But this is why a VAT would be the wrong policy. America’s fiscal problem is too much spending, not insufficient tax revenue. Imposing a new tax – particularly one capable of generating so much money – would be akin pouring gasoline on a fire and ensuring that America will become a European‐​style welfare state. In other words, bigger government and lower living standards.

Perversely, it is quite likely that a VAT would exacerbate rather than solve the problem of too much government borrowing. The nations in Europe that have been bailed out, as well as those teetering on the edge of fiscal collapse – including Greece, Spain, Portugal, Ireland, and Italy – all have VATs. Indeed, the average level of debt for all Western European nations is higher than the U.S. debt level. Imposing VATs, beginning in the 1960s, obviously didn’t keep politicians from spending too much. For all intents and purposes, the experience in Europe confirms Milton Friedman’s famous warning that, “In the long run government will spend whatever the tax system will raise, plus as much more as it can get away with.”

WASHINGTON -- The creation of a European-style Value Added Tax (VAT), which some policy makers are considering as a way to reduce the deficit, would come at a steep cost, according to an in-depth economic analysis commissioned by the National Retail Federation (NRF).

The report, which studied the effect of a 10.3% VAT applied on top of the current income tax system, says such a VAT would result in the loss of 850,000 jobs in the first year, reduce gross domestic product for three years, and bring a permanent drop in retail spending totaling $2.5 trillion over the first 10 years.

V. European VAT Changes A. Graphical Evidence First, we provide some graphical and nonparametric evidence by examin- ing the unconditional mean of the price index using the full sample of EU VAT changes in the 3 months before and after a VAT reform. In each case, we normalize the series to 100 in the month preceding the reform. Figure 7A plots the unweighted average price of all commodities sepa- rately for VAT increases and decreases. The figure shows that prices in- crease discontinuously in the month following a VAT increase but de- crease less in response to a VAT decrease. Importantly, this asymmetry is not driven by any subset of commodities. Indeed, when we plot disag- gregated versions of figure 7A by two-digit COICOP groups (in figs. E20– E22; figs. E11–E23 are available online), we find that all commodities exhibit this asymmetric pass-through, with the exception of communica- tion (COICOP group number 8) and furnishings, household equipment, and routine household maintenance (COICOP group number 5).19 B. Fixed-Effects Estimates To estimate the pass-through to prices of VAT increases and decreases, we first follow the approach of Evans, Ringel, and Stech (1999), who estimate the pass-through of cigarette taxes using different tax changes across US states over time. We run the following fixed-effects regression: D logðpict Þ 5 b0D logð1 1 tict Þ k510 (2) 1 o bkDlogð11ticðt1kÞÞ1Dlt 1gDXct 1Deict, k5210,k≠0 19 Possibly because of sample size, trends do not appear to be parallel for two COICOP categories: (1) alcoholic beverages, tobacco, and narcotics and (2) transport. While pass- through is asymmetric in both of these cases, the violation of the parallel assumptions sug- gest that we should be cautious in interpreting these two figures. 4456 journal of political economy FIG. 7.—Asymmetric response of prices to VAT changes. A, Response of the uncondi- tional means of prices to VAT increases and decreases. Lower panels plot the coefficients from the fixed-effects regression (2) for VAT increases (B) and VAT decreases (C ) on the full sample of reforms and include 10 month leads and lags. where i denotes the commodity, c the country, and t the month in which the price is observed, lt time fixed effects, pict the price, tict the VAT rate, and εict the error term. We control for a given country’s nominal interest rate, GDP per capita, and unemployment rate with Xct. For each of xt ∈ flogðpictÞ,logð11tictÞ,lt,Xct,eictg,Dxt isequaltoxt 2xt21. In equation (2), b0 identifies the pass-through of a VAT change in the month when the change occurs: for example, if b0 5 0, then the price does not respond to a VAT change, and if b0 5 1, the price responds one-to-one to a VAT change. The second term of the equation estimates any forward- or backward-looking responses of prices to changes in VAT rates; b25, for example, identifies the response of prices at time t to VAT changes that will occur at time t 1 5. The fixed-effects regression generalizes an event-study design with mul- tiple periods, commodities, and countries, and its main identification as- sumption is the same as that for difference-in-differences regressions: ab- sent the tax change, there would have been no change in the prices of the treated relative to the untreated commodities. Figure 7A shows a sharp change in prices at the time of the reform, with no pretrends and no evi- dence of anticipatory behavior, which lends support to this identification [chart omitted] what goes up may not come down 4457 assumption. Identification relies on within-country-specific commodity variation in VAT rates over time. In order to estimate the pass-through rate of VAT increases and de- creases, we estimate equation (2) separately on two samples. The first sample includes all observations treated by a VAT increase along with all untreated observations and is used to estimate the pass-through rate of VAT increases. Similarly, the second sample includes all observations treated by a VAT decrease along with all untreated observations and is used to estimate the pass-through rate of VAT decreases.20 Therefore, commodity-country pairs that never undergo a VAT change are included in both samples. The results of these fixed-effects regressions are reported in table 1.21 Columns (1) and (2) of table 1 correspond to VAT increases and decreases, respectively. The first row of each regression (labeled b0) shows the pass- through of the VAT change to prices 1 month after the reform. In partic- ular,b1i showstheresponseofpricestoVATchangesimonthsafterthere- form, while b2i shows the response of prices i months before the reform. Figures 7B and 7C plot the coefficients from regressions for VAT increases and decreases, respectively, and show that the pass-through to prices of VAT increases is equal to 34% while that of VAT decreases is equal to 6% one month after the reform; both are statistically significant. Importantly, there are no significant price responses in any months within a 10 month window around the VAT increases and decreases. We perform several ro- bustness checks, including running specification (2) without controls but with time fixed effects (table E8; tables E5–E23 are available online) separately on reforms that are classified as temporary and permanent (ta- bles E9 and E10), in periods of above- and below-median growth (ta- bles E11 and E12), on reforms that are not concurrent with other tax changes (table E13), with country-commodity specific inflation controls (table E14), and including more leads and lags (figs. E23A and E23B). We find similar levels of asymmetry in each case. In addition, we examine the relation- ship between the timing of VAT changes with underlying economic con- ditions and find that the two are not correlated. These correlations are discussed in detail in appendix B. 20 Untreated observations refer to commodity-country pairs that never undergo a VAT change as well as commodity-country pairs that are treated by a VAT change at a different time than the treatment period. 21 In a recent working paper, Abraham and Sun (2018) show that event study specifica- tions can be problematic under certain conditions. In our case, these concerns would ap- ply if there were heterogeneous effects across groups of VAT changes occurring at the same time. Given that our evidence is robust across a wide range of specifications as well as in several subsamples and also when plotting unconditional means in fig. 7A, this concern is unlikely to apply to our setting. 4458 journal of political economy TABLE 1 Pass-Through Estimates Using Fixed-Effects Regression (Full Sample) D log Price Increase Decrease (1) (2) .34 .064 (.063) (.030) .020 .025 (.019) (.017) .030 .026 (.019) (.015) .020 2.044 (.027) (.021) .015 2.005 (.016) (.028) 2.043 2.004 (.017) (.022) .049 2.020 (.034) (.022) 2.011 2.008 (.027) (.020) Yes Yes .014 .014 386,557 342,792 b0 b11 b22 b12 b23 b13 b24 b14 Time fixed effects R2 Observations Note.—The coefficients reported in this table show the pass-through of VAT increases and decreases to prices, estimated using specification (2) on the full sample of re- forms. Column 1 shows the estimates for VAT increases, and col. 2 those for VAT decreases. Standard errors are clustered by month and are in parentheses. b0 measures the pass-through of the VAT change at the time of the re- form, and bi measures price changes i months away from the reform. C. Matching Estimates In this section, we supplement our fixed-effects estimation with a match- ing approach that allows us to better mirror the setting of the Finnish ex- periment and to more directly interface with the theoretical predictions derived in section II. This approach also allows us to evaluate and alleviate concerns that in the preceding analysis we are never able to observe the effect of VAT increases and decreases at the same time, for the same com- modity, and in the same country. Our matching procedure follows Imbens (2015), and in particular that paper’s approach to analyzing the Imbens, Rubin, and Sacerdote (2001) lottery data (in sec. 6.1). First, and motivated by our theoretical framework from section II, we choose a set of key characteristics used to estimate pro- pensity scores. These characteristics include time, measures of economic activity, the magnitude of the VAT change, and the type of commodity what goes up may not come down 4459 TABLE 2 Pass-Through Estimates: Matched Sample With Trimming D log Price Increase Decrease (1) (2) .29 .042 (.10) (.047) Without Trimming D log Price b0 b11 b22 b12 b23 b13 b24 b14 Time fixed effects R2 Observations .021 (.019) .031 (.019) .021 (.027) .016 (.016) 2.042 (.018) .051 (.034) 2.0092 (.027) .025 (.017) .026 (.015) 2.044 (.021) 2.0050 (.028) 2.0047 (.022) 2.020 (.022) 2.0089 (.020) Increase (3) .33 (.066) .020 (.019) .030 (.019) .020 (.027) .015 (.016) 2.042 (.018) .049 (.034) 2.0096 (.027) Decrease (4) .056 (.037) .025 (.017) .026 (.015) 2.044 (.021) 2.0050 (.028) 2.0046 (.022) 2.020 (.022) 2.0089 (.020) Yes 384,559 342,453 386,080 342,477 Yes Yes .013 .014 .014 .014 Yes Note.—This table reports the pass-through estimates of VAT increases and decreases to prices estimated using specification (2) on matched reforms. Columns 1 and 2 implement the matching estimator on a trimmed sample using the approach outlined in Imbens (2015). Propensity scores are estimated using logit. Columns 3 and 4 report the estimates for the matched sample without trimming. VAT increases and decreases are matched on time, type of commodity (COICOP digit 2), size of VAT change, and GDP growth. In addi- tion, we match the preincrease VAT rate to the postdecrease VAT rate and the postincrease VAT rate to the predecrease VAT rate. Columns 1 and 3 show the estimates for VAT increases, and cols. 2 and 4 those for VAT decreases. Standard errors are clustered by month and are in parentheses. b0 measures the pass-through of the VAT change at the time of the reform, and bi measures price changes i months away from the reform. (two-digit COICOP code). Most importantly, guided by results from sec- tion II, we also match preincrease VAT rates to postdecrease VAT rates and predecrease VAT rates to postincrease VAT rates. Second, we use the es- timated propensity scores, based on these key characteristics, to trim the sample as suggested by Crump et al. (2009). Finally, in this trimmed sam- ple, we match the VAT increases and decreases, using the matching esti- mator outlined in section 5.4 of Imbens (2015).22 We then estimate equa- tion (2) on the matched sample of VAT changes. Estimates for pass-through from this matched sample, summarized in table 2, show that our finding of asymmetry is robust to matching. Indeed, 22 We cluster our standard errors as we do in our main specification. Bootstrapping yields similar standard errors, with all VAT increase pass-through estimates significant at least at the 1% level. However, we do not report these here, following cautions in Abadie and Imbens (2008) against using bootstrap with matching estimators. 4460 journal of political economy these estimates reveal asymmetries that are similar to those estimated us- ing the full (unmatched) sample of reforms. In appendix A, we also for- mally test the ability of our matching procedure to increase overlap in co- variates. We follow Imbens (2015), who uses normalized differences in order to assess overlap. Our matching procedure substantially improves overlap and reduces all normalized differences. While there is no formal threshold for normalized differences and while our matching procedure does not reduce all normalized differences to zero, they are well in line with those estimated by Imbens (2015) using the Imbens, Rubin, and Sa- cerdote (2001) lottery data. Because our matched sample mirrors the Finnish experiments, our pass-through estimates can be directly compared to the predictions of stan- dard incidence theory discussed in section II. We conclude that our find- ing of asymmetric pass-through in the matched sample is difficult to rec- oncile with static incidence theory. We provide more details on how the matching procedure is implemented in appendix A, and include formal tests of pre- and postmatching overlap in observables and further robustness checks. Note that a limitation of our approach is that we cannot match on prereform price levels because our data do not allow us to condition on price levels in a consistent way across all EU countries and commodities. This limitation implies that we are im- plicitly assuming that, conditional on matching on the key characteristics outlined above, any remaining variation in price levels has no meaningful effect over the asymmetry. For this reason, a stronger test of the asymmetry would also match on price levels. D. Long-Term Persistence of the Asymmetry In subsection IV.A we showed that asymmetry in the pass-through of VATs persisted for several years after the Finnish hairdressing reforms were en- acted. In this section, we provide evidence suggesting that this persistence is not a peculiarity of Finnish hairdressers: we observe it in other sectors and countries as well. In order to analyze longer-run horizons, it is impor- tant to focus attention on relatively large VAT changes and sectors in which prices are relatively stable. Absent these conditions, the effect of VAT changes on prices might be masked by natural variation in prices and in- flation that may be quantitatively significant in the long run. In order to study this we start by considering two case studies that in- volve large VAT changes. First, the European Commission approved an ap- plication to reclassify sit-down restaurants from the standard VAT rate to the reduced VAT rate.23 Both France and Finland took advantage of this 23 Following a campaign promise by then French president Jacques Chirac, France ap- plied for an authorization to reclassify sit-down restaurants from the standard to the re- duced VAT rate in 2002. The application was approved for all member states in January 2009. what goes up may not come down 4461 new law, leading to a 14 percentage point VAT cut for French sit-down res- taurants and a 9 percentage point cut for Finnish ones, in 2009. While the VAT rate did not revert to its original level as in the Finnish hairdressing experiment, we can exploit smaller increases in the reduced VAT rate: 1.5 and 3 percentage point increases in France, in 2012 and 2014, and a 1 percentage point increase in Finland, in 2013. Figures 8A and 8B show that the asymmetric pass-through persisted over several years both in Fin- land and in France. This evidence is to be interpreted cautiously, in part because the VAT changes are not of the same size and it is conceivable that the asymmetric pass-through is in part due to that. Second, and more comprehensively, Hungary cut its standard VAT rate from 25% to 20% in January 2006 and increased it from 20% to 25% in July 2009. These changes were enacted as part of a set of campaign prom- ises preceding the 2006 parliamentary elections. Figure 8C shows the re- sponse of commodities that were subject to the standard rate in Hungary FIG. 8.—Long-term persistence of the asymmetry. A, Response of Finnish sit-down res- taurants to a 9 percentage point VAT decrease and a 1 percentage point VAT increase com- pared to a control group of Norwegian sit-down restaurants. B, Response of French sit- down restaurants to a 14 percentage point VAT decrease and 1.5 and 3 percentage point VAT increases relative to a control group of Italian restaurants. We also include a hypothet- ical counterfactual that uses the VAT increase pass-through for VAT decreases. C, Response of all commodities subject to the standard VAT rate in Hungary (excluding diesel and gas- oline) to a 5 percentage point VAT decrease and a 5 percentage point VAT increase relative to a control group consisting of neighboring countries. D, Cumulative pass-through rates of all VAT increases and decreases in our sample over 20 months. [chart omitted] 4462 journal of political economy compared to a set of control countries.24 We find not only strong evidence of asymmetry but also that the asymmetry persisted over several years after the VAT rate was returned to 25%. Because the standard VAT rate applies to a wide range of commodities, this finding suggests that the long-term persistence of asymmetry may arise in more than a small set of specific sectors. Finally, we can assess the long-term persistence of the asymmetry in the full sample of reforms. To do so, we first reestimate equation (2) with 20 leads and 20 lags (we plot the resulting coefficients in figs. E23A and E23B). We find no substantial lagged responses, even 20 months after the changes take place. Second, we add up all of the lag coefficients and plot the cumulative pass-through rate for VAT increases and decreases over a 20 month period following the VAT changes in figure 8D. While both cu- mulative pass-through rates appear to decrease over time, the magnitude of the asymmetry—which is given by the difference between the two cu- mulative pass-through rates—remains relatively constant and shows no signs of convergence. However, in spite of the two time series being rela- tively stable, the precision of the cumulative pass-through estimates mechan- ically decreases over time, making any long-run statements past 18 months only suggestive.25 VI. Conclusion In this paper, we show that prices respond asymmetrically to VAT changes. First, prices respond more to VAT increases than to VAT decreases. Sec- ond, this asymmetric response results in an asymmetric pass-through of VAT changes to profits and markups. Third, the asymmetry persists over the long run. These patterns are inconsistent with the predictions of standard, static incidence models, and our findings therefore call into question the em- pirical relevance of these models for an important and widely employed tax. It is important for future research to examine how general these find- ings are and, in particular, to what degree the types of asymmetries we doc- ument arise with other taxes. More broadly, our findings suggest that it may be important to develop and empirically assess alternative models of incidence that can account for these types of asymmetries. One poten- tially promising avenue is to move beyond the static setting of classical 24 We included every commodity subject to the standard VAT rate, with the exception of diesel and gasoline because of strong volatility. Details of the list of commodities and con- trol group countries can be found in app. D (apps. C–E are available online). 25 We cannot reject that the cumulative VAT increase and decrease pass-through rates are equal after 18 months. We report the level difference between the two cumulative pass-through rates and the corresponding significance levels in table E15. what goes up may not come down 4463 models, and evaluate to what degree dynamic models of tax incidence can generate asymmetries like the ones we document.

Policy Recommendations All of the major tax systems implemented by governments must be seen as fair if they are to remain stable and sustainable in the long run. If a tax is viewed as unfair, then there will be constant demand to make adjustments to the system to address the demands of special interest groups. The problems associated with the administration of and compliance with a VAT system grow exponentially with the number of goods and services that receive tax preferences. In a single-stage sales tax system, a good is exempted from taxation, and only that good and its sales are affected in terms of special record keeping and additional compliance costs. However, with VAT, all sellers of such an item and the buyers who use it as an input in production must keep a special record of these sales and purchases separately. This is especially burdensome for small businesses that do not have sophisticated computer systems to track these special transactions. Even in the European Union today, small and medium enterprises account for more than 90% of the entire compliance costs associated with the VAT operation []. While in many countries, VAT has been the major tax instrument for raising revenue for their governments, it has in some cases collapsed. The experiences of Malaysia and previously of Ghana and Belize are examples. Usually, the collapse comes as a result of political concerns of tax fairness and compliance issues. In other countries, particularly in Latin America, it has occasionally been severely weakened by the proliferation of tax preferences and only salvaged by implementing major tax reforms []. Belize is a good case study of a situation where the collapse of the GST system seems imminent unless a major reform is undertaken. Over 40% of the entire workforce in GST tax administration are assigned to auditing and processing claims for refunds. Furthermore, the backlog of unprocessed claims in 2016 was equal in value to 50% of the annual revenues of the GST system. This is an institutionally unsustainable situation. This empirical study shows that the objective of raising a substantial amount of revenue with a single-rate, broad-based GST system in Belize is not in conflict with the government’s income distribution objectives. Our analysis shows that currently, the GST is highly progressive in Belize, with the effective tax rate paid by households in the top quintile 1.5 times the effective tax rate paid by those in the bottom quintile. However, the current tax system has required an extensive array of tax exemptions and zero rating of items, causing it to become impossible to efficiently administer. Even after eliminating the zero rating of all goods sold domestically and a drastic reduction in the number of GST-exempt goods, the tax system remains progressive. The average tax rate of households in the top quintile is approximately 1.2 times that of those in the bottom quintile.

Those who opposed Trump’s tariffs have a stronger position: China will not acquiesce in the U.S. imposing tariffs. Indeed China did not acquiesce. It retaliated against the United States by putting tariffs and, more importantly, administrative restrictions on imports of U.S. goods, particularly those pro-duced in states important to Trump’s political standing, including agricultur-

In one accounting Trump’s tariff war ended in an agreement with China within two years. Who won? It is hard for an objective observer to say, and likely the world is lucky that the economic warfare did not escalate.Starting early in the post-Second World War period as the frozen world thawed economists analyzed tariff warfare, a curse of the inter-war period (see Johnson, 1951, 1953, 1965). This literature strongly suggests that mov-ing from a free-trade equilibrium to a particular country’s optimum tariff provokes retaliation and ends in tariff warfare. Further both sides may well end with reduced welfare relative to free trade. Under some conditions, how-ever, one country may be better off, a boon to protectionists. Later analyses include a variety of complications.24 Games may be non-cooperative or co-operative, for one period or sequential, with or without signaling of various types or side payments or threats of various type. There may be interna-tional institutions to facilitate freer trade. More than two countries may be involved. Domestic special interests may represent various industries with influence. One can construct games in which a country may benefit by mov-ing from free trade to an optimum tariff even with the possibility of retalia-tion. History suggests the burden is on protectionists to provide a convinc-ing argument that a given proposed tariff is different from the base case of both countries losing.No country has trade as its sole international concern and this greatly complicates analysis. China is a difficult case for the U.S. free traders. China is an aggressive political, military and ideological competitor, not subtle and prone to insults, with much human-rights’ baggage. China has engaged in in-tellectual espionage and continues to do so. American politicians seldom lose ground with their voters by calling out the Chinese government and Chinese companies. Moreover, many Chinese leaders view free trade as a mere use-ful pretense. Giving China “one in the eye” sells politically.

China-US economic relations have expanded substantially over the past three decades, their mutual total merchandise trade rose from $2 billion in 1979 to $579 billion in 2016. China now is the US’ second-largest merchandise trading partner, third largest export market, and biggest source of imports. China-US mutual trade and investment relations benefit both countries, the US provides China a big exporting market, and the US imports of lower-cost goods from China greatly benefit US consumers. China is also the second-largest foreign holders of US Treasury securities, which helps US to keep interest rates low (Morrison ).

Despite growing economic relations and mutual dependence, China and the US bilateral disputes have become increasingly intensive. China criticizes the US of their export restrictions on high-technology products, their unfair treatments of China’s market economy status, and unreasonable trade sanctions on China. Major areas of concern expressed by the US include large numbers of trade surplus, relatively ineffective record of enforcing intellectual property rights (IPR), discriminatory innovation policies, and mixed record on implementing WTO obligations. Although faced with gradually increasing commercial disputes, the China–US economic relations are positive and develop fast in general. But the harmonious situation seems to change after Donald Trump was elected to be the President of the US

To quantify the degree of tariff changes that firms faced, we review the tariff changes implemented under the Trump administration up to the point where the trade war was put on pause by a Phase one agreement.

Immediately after taking office, the Trump administration dramatically reversed the longstanding U.S. course towards  by launching numerous actions which imposed new tariffs on named partners. The initial Trump tariffs, which were rationalized by argued threats to national security led to U.S. global safeguard tariffs on imports of washing machines and solar panels in January 2018, and tariffs on steel and aluminum imports in March 2018. The initial tariffs were focused on a few products and applied to all trading partners (with a few exceptions), and eventually led to retaliatory tariff action by some countries including China, Canada, Mexico, and the European Union.

The ratcheting up of tariffs and the Fed’s discretionary conduct of monetary policy are a toxic mix for economic performance. Escalating tariffs and President Trump’s erratic and unpredictable trade policy and threats are harming global economic performance, distorting monetary policy, and undermining the Fed’s credibility and independence.

President Trump’s objectives to force China to open access to its markets for international trade, reduce capital controls, modify unfair treatment of intellectual property, and address cybersecurity issues and other U.S. national security issues are laudable goals with sizable benefits. However, the costs of escalating tariffs are mounting, and the tactic of relying exclusively on barriers to trade and protectionism is misguided and potentially dangerous.

However, unfortunately, neither policymakers in the U.S. nor in the EU have so far publicly considered an additional policy measure which could solve several problems at once: to revive and complete an EU-U.S. free trade agreement, building on the Obama administration’s efforts to establish the Transatlantic Trade and Investment Partnership (TTIP). A TTIP successor would not only have the potential to significantly ease existing supply chain woes and reduce inflation in the medium-term but also increase economic growth on both sides of the Atlantic, thereby reducing the risk of a prolonged recession in the U.S. and the EU. While reviving an EU-U.S. free trade agreement will take some time, central bankers now expect that inflation pressures might last longer than expected and that we might not be returning to the low-inflation environment we faced before the pandemic. This underscores the importance of finding other policy measures that curb inflation but do not reduce economic growth. Moreover, an EU-U.S. free trade agreement would lead to greater integration between these two vital blocs, deepening economic and political ties in line with the existing close military ties, increasing regulatory harmonization, and building a more coherent counterweight to the increasingly assertive authoritarian alliance between Russia and China. Reviving a TTIP successor could help improve the economy, reduce inflation, and strengthen ties with our democratic allies.

President Biden would have a good basis to restart negotiations with the EU by building on the far-reaching efforts to implement TTIP by the Obama administration. Before the Trump administration scuttled TTIP and antagonized the EU by imposing tariffs on European goods, President Obama had made significant progress in negotiating an EU-U.S. free trade agreement after launching talks in 2013. The U.S. could pick up where it left off in 2016 by adopting the provisions settled by the initial 15 rounds of negotiations and renewing these talks.

A specter is haunting the international economy: the specter of a trade war. Well, the specter of a trade war is at least haunting economists. It is striking that the distaste for protectionism felt by the discipline is not shared by the wider public. Modern economics began over two hundred years ago in part as an intellectual exercise against mercantilism, so it is worrying that the profession has been unable to persuade the public of the merits of free trade. But perhaps some of the public’s mild views on protectionism stem from the fact that most economic analysis of protectionism is theoretical, microeconomic, or dated? In this paper, we examine the macroeconomic consequence of tariffs. We use impulse response functions from local projections on a panel of annual data spanning 151 countries over 1963-2014. The main analysis on aggregate data is complemented with industry-level data. Our results suggest that tariff increases have an adverse impact on output and productivity; these effects are economically and statistically significant. They are magnified when tariffs are used during expansions, for advanced economies, and when tariffs go up. We also find that that tariff increases lead to more unemployment and higher inequality, further adding to the deadweight losses of tariffs. Tariffs have only small effects on the trade balance though, in part because they induce offsetting exchange rate appreciations. Finally, protectionism also leads to a decline in consumption; this, together with our other findings, suggests that tariffs are bad for welfare. All this seems eminently sensible and bolsters the arguments that mainstream economists make against tariffs; our results can be regarded as strong empirical evidence for the benefits of liberal trade. And given the current global context, we take special note of the negative consequences when advanced economies increase tariffs during cyclical upturns.

This study attempted to estimate the effect of a change in labour taxation in Brazil on employment and wages. Starting in January 2012, firms not under the special tax regime for small and micro enterprises (Simples) that operated in four economic sectors experienced a cut in the employers’ contribution to social security of 20 percent of the wage bill. Relying on administrative data that allowed for a firm level analysis, the study explored information on the tax regime to build a potentially credible counterfactual necessary for the measurement of the policy effects. The main results suggest that treated firms increased the number of formal employment contracts by 14-16 percent, total working hours by 8-10 percent, and wages by 2-2.3 percent, in relation to untreated firms in the same sector and size range. These figures indicate that the payroll tax reduction resulted in employment generation, increase in labour demand, and in a small increase in wages in its first year of implementation. These estimates are statistically significant in almost all specifications. They seem robust to the exclusion of the textile sector, which was only covered later, as the effects become stronger if the analysis is limited to firms in the sectors covered earlier. The study also confirmed that employment and labour demand were not anticipated by firms in 2011, as the bulk of the policy effects seem to have happened after the policy adoption. Besides, weaker effects on employment and labour input were found when larger firms are included in the analysis. When the analysis is disaggregated by sector, the largest coefficients of variation in employment and wages were estimated in the leather and shoes industry, while the textile sector shows the smallest. The garment industry and services sectors fell in between the two. On the one hand, the leather and shoes industry is the most export-oriented sector, thus may have captured the full benefit of a payroll tax reduction designed to stimulate exports. On the other, the textile sector was only fully covered by the policy at a latter moment in time, which seems to explain the smaller effects in 2012. Empirical studies carried out in Latin-America and elsewhere tend to show small employment effects and partial shifting of payroll tax changes on to wages (Gruber 1997, Cruces et al. 2010, Bennmarker et al. 2009). This study points in the opposite direction, suggesting a substantial employment effect. The reasons might be that the analysis here is limited to a few sectors and does not attempt to evaluate the overall impact on the aggregate employment level. The employment effect is consistent with the theory that a decrease in labour costs results in an expansion in the demand for labour at firm or sector levels. The shift in demand interacts with an elastic labour supply to the formal labour market, determined by the size of the informal economy and the high unemployment rate, resulting in the adjustment by the employment with smaller effect on wages. 48 Results presented here contrast with two studies about the Brazilian payroll tax reform that found small positive effects on employment, though not across all covered sectors (Dallava 2014, FGV Projetos 2013). It is believed that the empirical approach adopted in this paper, and the control for firm fixed effects, offers a better solution to the identification problem that comes from the evaluation of such policy, as seen by more intuitive results and highly significant estimates. Once data for 2013 and 2014 are available, a broader policy evaluation using this approach could include sectors that were covered in the second half of 2012 onwards. It is also possible to undertake further evaluations using other data sources that contain information about relevant covariates such as capital intensity, and gross revenue. Moreover, the aggregate impact of the payroll tax reduction and its substitution by a revenue tax could be evaluated more broadly by use of an adequate methodology and including fiscal and inter-sectorial interactions. The policy seems to have achieved the expected increase in formal employment and perhaps an enhanced competitiveness, at least for the covered industries and in its first year. However, given the limitations of this study, these conclusions do not imply that aggregate effects will be the same. Besides, reductions in other non-wage labour costs in Brazil may not have similar effects, as changes in each component of labour costs will trigger specific effects on the labour market. Instead of a simplistic generalization, it seems also necessary to consider the implications of changes in social security, training, and the financing of other social programmes

Conclusion: This paper develops a general theory of state-dependent fiscal multipliers for a broad range of spending and taxation policies. The framework accounts for empirically relevant goods market frictions by incorporating idle productive capacity and unsatisfied households’ demand into an otherwise standard general equilibrium setup. Our key novel finding is that cyclicality of fiscal multipliers is pinned down by the source of economic fluctuations, and we provide model-free econometric evidence that strongly supports our predictions. Crucially, we establish that multipliers associated with fiscal instruments which stimulate aggregate demand, such as government spending and consumption tax cuts, are countercyclical under demand-driven fluctuations and procyclical under supply-driven fluctuations. On the other hand, multipliers associated with interventions that stimulate aggregate supply, such as reductions in taxes on firms’ payroll, sales and households’ labor income, are countercyclical under supply-driven fluctuations and procyclical under demand-driven fluctuations. In addition, our theoretical results establish a relevant role for fiscal austerity, implemented by a reduction in government consumption in severe supply-driven recessions and demand-driven booms, provided elasticities of labor supply and labor demand are sufficiently low. Further, we develop and estimate a novel econometric specification that allows us to perform model-free evaluation of both spending and taxation multipliers in recessionary and expansionary episodes, conditional on those being either demand- or supply-driven in nature. Our empirical results detect substantial state dependence, conditional on the source of fluctuations, which is in line with the predictions of our theory. Such findings offer a resolution to the debate on state dependence of fiscal policy, on both empirical and theoretical grounds, and they provide guidance for the conduct of fiscal policy in the different phases of the business cycle. Our analysis opens fruitful avenues for future research. First, in our dynamic framework, current changes in fiscal policy determine the future path of goods market tightness and thus constrain the effectiveness of policy in the future. Our framework can therefore be extended to study the intertemporal trade-offs and the path dependence of fiscal policy. Second, by extending the model to include heterogeneity in the goods markets, one can study how composition of spending and taxation policies may generate of fiscal policy across sectors and the socially optimal distribution of such policies. We plan to investigate these issues in future research.

Conclusion: This paper shows that personal income taxation is negatively related to corporate investment. The magnitude of the investment responses to personal tax changes at lower income levels is comparable to that of the investment responses to corporate tax changes in our sample. It further shows that the PIT–investment relation can be explained by two channels. First, higher PIT increase labor costs, which results in stronger investment responses by firms facing more elastic workers, by financially constrained firms, and by firms in industries with a strong link between capital and labor input. Second, PIT can affect investment through reduced consumption. Our findings have important implications for the debate on tax policy design, since PIT are a major revenue source for governments. Our results indicate that personal taxes can have a substantial effect on investment, particularly PIT at lower income levels. Hence, our results provide important input for tax policy considerations. The reduced magnitude of the effects at higher income levels suggests that governments could balance their budgets while avoiding negative investment effects by lowering taxes at lower income levels and financing the shortfall by implementing modest tax increases at higher income levels. Of course, we do not provide a full welfare analysis. Still, given that investment is a key driver of overall economic growth, the documented investment responses appear to be important for academics and, in particular, policymakers, as policymakers facing budget deficits may consider PIT increases to balance their budgets.