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The Determinants of Cross Border Shopping

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National Tax JournalVol. LIII, No. 4, Part 1

Abstract - This paper analyzes the choice by Canadian consumerswhether to cross the border into the U.S. to shop. To do so a modelis built in which consumers value two consumption goods (goodsthat can and cannot be smuggled), leisure, and government ser-vices (provided through commodity taxes). The model’s predictionsare tested against same–day border crossing data for the period1972:01 through 1997:12. The results are then used to estimate thetax revenues forgone from the introduction of the GST in Canada.The data also suggest an extension in our thinking about the tradi-tional domain of policy responses—from the use of alternative taxesto institutional and/or regulatory change.

INTRODUCTION

In this paper I focus on the determinants of the decision byindividuals to cross the border to shop and test the predic-

tions of the model developed for that purpose on Canadianmonthly border crossing data for the period running between1972 and 1997. For economists and policy makers, crossborder travel is of particular interest because of its closeassociation with sales tax evasion and, as such, this paperforms part of the growing literature on international integra-tion and its impact on tax structure (see Kanbur and Keen,1993). Most often mobility (whether it is product, capital,or labor mobility) is seen as a byproduct of or constraint onthe game theoretic fiscal competition that arises among theregions of a single state and hence is most often tested oninterstate or provincial mobility (e.g., Mintz and Tulkens,1986; Trandel, 1992, 1994; and Thursby, Jensen, and Thursby,1991). Here I utilize the extensive data available on interna-tional border crossings to test a shopping model that hasimplications for tax evasion through the return flow ofshoppers.

The motivation for this paper arises from a brief consider-ation of the set of policy choices faced by Canada in the pe-riod leading into and immediately following the Free TradeAgreement with the United States in January, 1989. With theprospect of fewer barriers to the movement of both goods

The Determinants of Cross BorderShopping: Implications for Tax Revenues

and Institutional Change

J. Stephen FerrisDepartment ofEconomics,Carleton University,Ottawa, Canada

NATIONAL TAX JOURNAL

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and factors across the Canada–U.S. bor-der, policy makers entered a new environ-ment where international factor mobilitymade problematic traditional responses tochanges in tax levels and tax instruments.In particular, Canadian policy makersappeared to recognize that continued re-liance on a relatively high manufacturer’ssales tax would likely reduce the level ofmanufacturing in Canada and lead to itsrelocation into the United States. This be-gan a period of iterative policy responseas Canada abandoned the use of a high,narrow producer tax (the manufacturer’ssales tax) and adopted a lower, more gen-eral value added tax (the Goods and Ser-vices Tax, GST). Some of the behavioralresponses to the new policy regime forcross border shopping have been recog-nized already. Di Matteo (1993), for ex-ample, has documented the higher flowof cross border travel that followed theintroduction of the GST and more recenttheoretical analysis has focused on thegain in private consumer utility from crossborder tax evasion and questionedwhether net social gains can be producedby tighter enforcement of border crossingregulations (Lovely, 1994, 1995). Myanalysis focuses on the broader determi-nants of cross border shopping of whichtax evasion represents only one dimen-sion.

At bottom, however, a policy that raisesthe benefit or reduces the cost of crossingthe border to evade taxes has social sig-nificance because it limits the revenue thatcan be raised from the current tax struc-ture and hence increases the cost of pro-viding goods and services through thegovernment. Moreover, the distortion cre-ated by the misperception that individu-als can escape taxation by cross bordershopping can be expected to prompt fur-ther change as policy makers respond tothe altered cost of using traditional taxesand maintaining the same regulatory en-vironment. As we will see, Canada ap-

pears to have used a wider range of re-sponse to trading openness than wouldusually have been considered. Of particu-lar interest is the suggestion in the datathat shopping regulations responded tothe incentives created by the opportunityto evade taxes that arise in more openmarkets.

The analysis begins by adopting amodel developed by Lovely (1994) to ana-lyze cross border tax evasion and enforce-ment alternatives. Rather than use themodel to examine the welfare issuesraised by cross border shopping, themodel is used positively to predict thedeterminants of cross border shopping.Next the comparative static effects on thenumber of cross border shopping tripsfollowing changes in the costs and ben-efits of cross border shopping are pre-sented. The next section describes the dataused in the tests and is followed by theempirical results. The equation estimatesare then used to measure the size of theexpected tax losses that follow the intro-duction of the GST in Canada and thisestimate is compared with other evidenceon the aggregate size of the tax evasiontax loss. The paper concludes with lessonslearned.

THE BASIC MODEL

Consider an open economy with alarge number of identical price–takingconsumers (normalized to one for conve-nience). Consumers are assumed to treatsimilar goods identically even if producedin a different country and thus see them-selves as being able to purchase goodsfrom two sources: an internal retailmarket (that sells both domestic goodsand legal imports) and an external mar-ket (where foreign produced goodscan bepurchased but must be smuggled in fordomestic consumption). To distinguishbetween types of goods, I denote goodsthat cannot be smuggledwith the sub-


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script n, while goods that can be smuggledare given the subscript s.1 Superscripts areused to indicate their production originas either domestic, d, or foreign, f. Usingthese distinctions, aggregate domesticconsumption will consist of both types ofgoods, Cs and Cn. Nonsmuggled consump-tion goods are produced domestically (sothat Cn = Cn

d), while goods that can besmuggled are either produced domesti-cally, Cd

s, or in the foreign country, Csf.

Because crossing the border to shop iscostly in terms of time lost from work orleisure, individuals will prefer to shop indomestic retail stores (ceteris paribus). Onoccasion, however, cross border shoppingbecomes desirable and the extra time andeffort required to cross the border adds atravel cost that is largely independent ofthe number of purchases. This means thatcross border shopping will take place dis-cretely in time and suggests that crossborder shopping can be metered by thenumber of shopping trips, n.2 Becausecross border shopping typically involvesthe use of a car, the quantity and type ofgoods that can be safely brought backwithout detection each trip is strictly lim-ited. Using cs

f to denote the bundle ofgoods purchased in a representative crossborder shopping trip, Cs

f = ncsf. In the

model below, csf is taken as a parameter

whose size implies n > 1. This assump-tion is more fully motivated in the datasection below.

The consumer’s shopping decision ismodeled by assuming that the represen-tative consumer maximizes a utility func-tion that is strictly concave with continu-ous first and second derivatives in the

arguments: consumption, leisure, andgovernment services. That is:

[1] U = U(Cn, Cs, l; g)

where the C’s represent units of the twoconsumption goods; l represents hours ofleisure; and g represents the level of gov-ernment spending (assumed exogenous toeach consumer). The latter assumption in-troduces a fiscal externality into the modelin that individuals believe that the govern-ment services they receive are independentof their consumption and work decisions(despite being linked directly through thegovernment budget constraint).

Each consumer is constrained by bothincome and time. The income constraintof the representative individual is writtenin units of the domestic currency as:

[2] (1 + t)pnCn + (1 + t)psCsd + eps

fncsf = wh,

where e is the domestic price of foreigncurrency (number of Canadian dollars perU.S. dollar); ps and pn represent, respec-tively, the domestic price of goods that canand cannot be smuggled; ps

f is the price ofsmuggled goods in the foreign currency;w is the hourly wage; h is the number ofhours worked; and t is the common do-mestic commodity tax rate.

The second constraint accounts for timeand is represented as:

[3] l = H – h – nδ.

Here H represents the maximum numberof hours available; h, the number of hoursspent in work; n, the number of cross

1 Smuggled goods, as used in the text, fall into one of two categories: goods brought in legally through narrow(e.g., art) or broad (e.g., all goods purchased within duty free limits) exemptions in custom duties and goodsbrought in illegally. Same–day cross border shopping, the focus of this analysis, has no duty free exemptionso that the method of bringing in goods (usually by car) and the necessity of hiding undeclared importsdefines the subset of goods that can be smuggled and the use of trips as a meter of smuggled quantities.

2 Kanbur and Keen (1993) locate households across space so that distant households face a higher cost of cross-ing the border to purchase a fixed quantity of goods. My analysis focuses on a representative household whopurchases a variable quantity of goods by changing the number of shopping trips. In this sense my analysisfocuses on the intensive margin while Kanbur and Keen focus on the extensive margin.


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border shopping trips; and δ, the hoursspent on each cross border shopping trip.

Finally, g is used to represent units of apublic good that can be produced at a con-stant cost, cg. Public goods are financed bya common commodity tax rate, t, leviedon all goods sold domestically. Normal-izing the measure of government servicesto produce a marginal cost equal to onethe government budget constraint be-comes:

[4] g = t(pnCn + psCsd).

Government spending is set exogenouslyso that the tax rate becomes a residual inthis analysis. The tax rate then adjustsautomatically so that the collectionsneeded to finance g are consistent with thelevel and composition of private con-sumption.3

Substituting the income and time con-straints of equations [2] and [3] directlyinto the utility function eliminates twoof the interdependent choice variables(Cn

d and l) from explicit analysis.4 Aftersubstitution the objective function be-comes quasi–concave and a maximum canstill be found. Writing this out explicitly,the maximization problem becomes:5

[5] Max U(Csd , n, h) = U[(w/(1 + t)pn)h

– (ps/pn)Csd – (eps

f /(1 + t)pn)ncsf, Cs

d

+ ncsf, H – h – nδ; g].

Equation [5] formalizes the case where theindividual does not recognize the linkbetween tax collections and governmentservices and believes that he or she can

free ride on the provision of governmentservices by avoiding tax paymentsthrough cross border shopping.

The first order conditions for the inter-nal optimum when the individual buysthe good that can be smuggled in both theforeign country and domestic market are:

[6] Csd: UCs/UCn = ps/pn,

[7] n: (1 – epsf/(1 + t)ps)cs

f UCs = δUl,

[8] h: Ul = (w/(1 + t)pn)UCn,

where [6] has been used in [7] to allow amore intuitive interpretation. From [7]note that n > 0 requires 1 – eps

f/(1 + t)ps >0; that is, the full price of the domesticgood that can be smuggled, (1 + t)ps, mustexceed the domestic price of the foreigngood, eps

f, for cross border shopping to be-come economic. An interior solution canthen arise in which the individual pur-chases goods that can be smuggled in boththe domestic and foreign market becauseof the assumption that cross border shop-ping is relatively time intensive and theutility function exhibits diminishing re-turns to leisure.

The first order conditions describe thethree margins of choice that form the ba-sis for the model’s predictions. Given theinterior solution, equation [6] describesthe optimal consumption choice acrossdomestic good types as one where the in-dividual will increase consumption of thedomestic good that can be smuggled aslong as the net utility gain exceeds the lossarising from the reduced consumption ofgoods that cannot be smuggled. In thedomestic market, both goods are subject

3 The addition of a theory of government allows analysis of the welfare implications of the model. Hence aLeviathan theory will generate a level of government that is too large (relative to individual preferences)while a model with free riding hypothesis generates a level too small.

4 Recognizing that Cs = C

sd + nc

sf allows rewriting the budget constraint to solve for C

nd as C

n = [w/(1 + t)p

n]h – (p

s/

pn)C

sd – [ep

sf/(1 + t)p

n]nc

sf. Substituting the government budget constraint directly into the utility function

would force recognition of the consequences of individual behaviour on the level of government services,internalizing the fiscal externality.

5 After substituting for Cnd in g, the two C

sd terms cancel.


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to the same tax rate. Equation [7] describesthe condition for the optimal number ofcross border shopping trips as one wherethe individual continues to cross the bor-der and shop as long as the net utility re-ceived by purchasing the set of foreigngoods at a lower price (in domestic cur-rency) exceeds the utility cost of foregoneleisure. Finally, equation [8] establishes theoptimal labour/leisure choice by balancingthe loss in utility from working longer hours(lost leisure) against the gain from higherincome and consumption (in terms of themarginal utility of goods not smuggled). Inmaking this tradeoff, the individual doesnot include the net utility gain that arisesfrom the additional government servicesarising from higher realized taxes.

Equations [6], [7] and [8], together withthe government budget constraint in [4]and a rule for determining governmentsize, are sufficient to solve for the repre-sentative individual’s optimal tradingplan. If the country is a small openeconomy (like Canada), these conditionsare sufficient to describe the choices made.

COMPARATIVE STATICS

In this section I derive the predictionsthat are tested in the empirical analysis. Inparticular, equations [6], [7], and [8] areused to solve for the change in the numberof cross border shopping trips that followfrom changes in: the foreign price of goodsthat can be smuggled, ps

f; the domestic com-modity tax rate (price), t (ps); the exchangerate, e; the quantity of goods that can befeasibly smuggled on each shopping trip,cs

f; the time required to travel, δ; and thewage rate, w. While most of these variableshave a straightforward counterpart in thedata, changes in the time cost of shopping

abroad instead of at home are not directlyobservable. This led me to considerchanges in institutional arrangements thatmight be expected to affect border cross-ing times as proxies for δ. For example, theU.S./Canada Free Trade Agreement(USFTA) beginning in January, 1989 wouldbe expected to reduce the time and admin-istrative ease of crossing the border andhence correspond to a fall in δ in the model.

Formal presentation of the comparativestatic results are available from the authoron request. Here I simply report the quali-tative model predictions as: dn/dps

f < 0; dn/dt > 0; dn/de < 0; dn/dcs

f > 0; dn/dδ < 0; anddn/dw < 0. The first two terms reflectchanges in the relative price of goods thatcan be smuggled and the effect this is ex-pected to have on cross border shopping.An increase in the foreign price raises therelative price directly while an increase inthe domestic commodity tax rate raises thedomestic price, so lowering the relativeprice of the foreign good. In either case, anincrease in the relative price of foreigngoods will reduce the value of cross bor-der shopping and hence reduce the num-ber of cross border trips. These two changeswork as described only if the exchange rateis held constant. Offsetting movements inthe exchange rate could undo the relativechange in cross–country prices. More gen-erally, any increase in the exchange rate (i.e.,an increase in the number of Canadian dol-lars needed to acquire a U. S. dollar) willraise the domestic cost of acquiring thesame good in the U.S. and so lead to a fallin Canadian cross border shopping. On theother hand, an increase in the number offoreign goods that can be brought back eas-ily on each shopping trip without payingthe domestic tax is expected to increase thenumber of cross border shopping trips.6

6 Note again that this result reflects the assumption that border enforcement limits “artificially” the number ofpurchases made each trip so that smuggling remains intramarginal. In this case raised exemptions, less en-forcement, and/or newer smuggling techniques allow lower cost foreign goods to replace higher cost domes-tic goods so that the marginal value of an additional trip comes to exceed the higher time cost of cross bordershopping. The opposite result would arise if cross border shoppers exhaust the desired commodity margin sothat fewer trips are now needed to purchase the same quantity of goods.


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Lastly, any increase in the time required tocross the border and shop (relative to do-mestic alternatives) or an increase in thevalue of that time (as represented by a risein the wage rate) is predicted to decreasecross border shopping. These predictionsare tested and the size of the estimated ef-fects quantified below.

THE DATA

The major difficulty associated withtesting the shopping model is that the ac-tivity of crossing the border to evade com-modity taxes is illegal and must be in-ferred from data collected for other pur-poses. Hence the test below focuses on thenumber of individuals crossing the bor-der (rather than the shopping trip) and theregression analysis uses same–day cross-ings by automobile [AUTOS] and by in-dividuals in automobiles [SAMEDAY] asa proxy for cross border shopping trips.While there are many reasons for cross-ing the border, same–day shopping tripsto Burlington or Massena (from Montreal),Watertown (from Ottawa and Kingston),Buffalo (from Toronto), Detroit (fromWindsor), and Bellingham (fromVancouver) have long been used by Ca-nadians to stock up on relatively highpriced Canadian goods and for this rea-son same–day crossings are often used asthe barometer of the scale of cross bordershopping activity for policy analysis (e.g.,

Di Matteo, 1993).7 However, while crossborder travel may accurately measure themovement of consumers to foreign goods,commodity taxes can also be evaded byhaving goods brought to consumers byprofessional smugglers. The absence ofreliable data on changes in organizedsmuggling flows can then bias an analy-sis that measures evasion solely by focus-ing on individual crossings. For example,in late 1993 and early 1994 organized to-bacco smuggling into Canada became soprevalent that the domestic retail marketfor tobacco in Quebec and Ontario wasvirtually eliminated. For this time period,the widespread availability of smuggledtobacco inside Canada reduced cross bor-der travel for tobacco products to a trickle.In this particular case, the large loss in taxrevenue and fear of the public’s growingacceptance of tobacco smuggling ledCanada’s governments to cooperate anddrastically drop (in February, 1994) spe-cific federal and provincial tobacco taxesto remove the incentive to smuggle.8

While such episodes remain the exceptionrather than the rule, it is important to re-member that same–day travel is simply ameter rather than a precise measure of theextent of tax evasion.9

Figure 1 presents the monthly time se-ries of individual same–day border cross-ings from Canada to the U.S. betweenJanuary, 1972 and December, 1997 (312observations). I also include on that dia-

7 Same–day travel has no federal import duty exemption (while increasing exemptions are given after twenty–four hours, forty–eight hours, and one week abroad) so that all goods purchased outside should be declaredfor the payment of the appropriate federal and provincial taxes. Despite this requirement, considerable num-bers of Canadians have at one time or another crossed the border to engage in same–day shopping withoutdeclaring their purchases.

8 Early empirical work for the province of Quebec showed no same–day crossing response to tax inclusivechanges in cigarette prices despite Quebec’s growing tax differential with the U.S., a relatively large smokingpopulation, and a short travel distance to the U.S. This became more understandable once it was recognizedthat by 1993 the scale of organized cigarette smuggling had made personal travel redundant. Note, however,that Quebec remains an anomaly in relation to its response to extended opening hours (see the regressionequations below).

9 The positive aspect of border crossing data is that because governments do monitor their borders, the datacollected does allow particular focus on the trip. That is, domestic wholesale and retail data typically relate tostores or products, while most shopping hypotheses emphasize the time cost of shopping and so generatepredictions in terms of numbers of trips.


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gram the exchange rate (measured as thenumber of Canadian dollars per U.S. dol-lar). This illustrates that more than sim-ply the exchange rate is needed to explaincross border travel. Perhaps what is mostapparent from the same–day series is itsrepetitive seasonal pattern. Of greater sig-nificance for the shopping hypothesis,however, is the dramatic rise and fall intravel that takes place towards the end ofthe time period. Between 1987 and 1992,same–day cross border travel more thandoubled and even by the end of the timeperiod in 1997, same–day crossings wererunning at a rate that was 50 percenthigher than at the beginning. What is lessclear from the figure is that this time se-ries (SAMEDAY) is also nonstationary.10

To achieve stationarity (required for sta-tistical testing), the series was firstdifferenced. Using the difference opera-tor D to represent first differences, the testspresented below use D(SAMEDAY) astheir primary dependent variable. To testfor the robustness of the “individual”measure of cross border shopping, how-ever, I also used the number of automo-biles making same–day crossings (AU-TOS). AUTOS shared the time seriescharacteristics of SAMEDAY and so werefirst differenced to generate the alterna-tive dependent variable D(AUTOS).11

Testing the shopping hypothesis in firstdifference form is consistent with the timeseries properties of the other key indepen-dent variables of the analysis (i.e., the rela-

Figure 1. Same–day Border Crossings: 1972:01–1997:12

10 The adjusted Dickey–Fuller (ADF) test statistic for SAMEDAY is –3.08, below (in absolute terms) the MacKinnoncritical value of –3.45 at 1 percent and only marginally above the 5 percent critical value of –2.87. In firstdifferences, the ADF statistic becomes –8.703, which is consistent with stationarity.

11 The ADF test statistics for AUTOS are –2.141 in levels and –8.815 in first differences.

Num

ber

of S

ame-

day

Cro

ssin

gs

Can. $ per U

.S. $


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tive price ratio and the exchange rate arealso integrated of order one).12 However,the use of first differences alone will leadto the loss of potential information thatmight be inferred from the relationshipamong the levels of the variables. Becausethe levels of a set of variables may becointegrated despite being individuallynonstationary, I tested for the stationarityof the residuals of the regression equationwhen run in levels. The possibility of anerror correction relationship was sug-gested by the ADF test statistics of the re-siduals of the SAMEDAY and AUTOSregression equations—taking the values–4.95 and –7.42, respectively, both ofwhich exceed (in absolute terms) the1 percent critical MacKinnon value (–3.45).An error correction formulation then addsto the first difference form of the regres-sion equation a term to reflect the laggederror term of the same equation when runin levels (the error correction term). Thepredicted sign of the coefficient of the er-ror correction term is negative (indicatingconvergence). Its significance would sig-nal the presence of a convergent processby which individuals adjust their shop-ping trips to past deviations (errors) ofactual from desired levels of crossing. Asimportantly, the presence of the error cor-rection term corrects for bias in the esti-mated fit of a simple first difference equa-tion (by accounting for an otherwise un-explained adjustment process) and allowsfor more precision in the identification ofeach hypothesized relationship. The errorcorrection model used to test the predic-tions of the model below estimates thelonger run error correction term and the

shorter run first difference relationshipsimultaneously. This allows a direct esti-mate of both the short and long run re-sponse of cross border same–day travel toeach of the model’s key variables.

While higher commodity taxes may cre-ate a general incentive for Canadians tocross the border and shop, the particularcommodities purchased on that trip de-pend upon the ability of shoppers to bringthem back undetected. Because all unde-clared same–day retail purchases are ille-gal, only a subset of potentially profitablecommodities will be purchased and thesewill be chosen as much for their smug-gling characteristics as for the benefit gen-erated by commodity tax evasion.13 Thecharacteristics needed for smuggling arereadily apparent—small in size, high inrealizable savings, and low in visibility—and these criteria were used to pick a sub-set of potential commodities to measurethe dimension of the smuggling bundle.In addition, virtually all cross bordershopping is done by car and travel by carpermits a relatively small number of waysof avoiding superficial detection at theborder. It is this inability to smuggle morethan small quantities at relatively low costthat motivates our assumption that cs

f, thebundle size, is fixed. Without this risingcost of detection, increases in tax differ-ences could well result in larger purchasesper trip rather than a larger number oftrips.

Five particular commodity classes werechosen to meter the price incentive to crossborder shop: apparel, food, gasoline, to-bacco, and liquor. Apparel and gasolineare natural candidates for inclusion in the

12 The ratios of both exchange rates and unadjusted relative prices are all nonstationarity in their levels. TheADF test statistics all fall short of the critical MacKinnon value of –3.45 for levels and exceeded it for firstdifferences. The ADF values for EXCH and the relative price of the smuggling bundle (PRBUNDLE) are,respectively, –0.905 and –1.55 in levels and –7.91 and –7.38 in first differences.

13 While the analysis focuses on smuggling, the same factors are at work if differences in commodity taxes andimport duties allow for profitable declaration on return. Throughout the early part of our time period, aCanadian returning from the U.S. could bring back duty free: $20 worth of products after twenty-four hours;$100 after forty-eight hours; and $300 after seven days, with further restrictions on the quantity of tobacco andliquor that could be imported duty free. On June 13, 1995 these limits rose to $50, $200, and $500 (without,however, introducing exemptions for same–day travel).


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smuggling bundle since the scale of legiti-mate trade in these products makes thedetection of place of purchase virtuallyimpossible if done in small quantities.14

The period of construction of “U.S. fac-tory outlet” clothing stores adjacent to theCanadian border in the early 1990sseemed to be designed precisely to accom-modate this purpose.15 Food is present inthe list not for sales tax reasons (food prod-ucts being exempted from sales taxes inCanada) but because Canada relies to agreater degree than does the U.S. on mar-keting boards to control domestic agricul-tural supplies. This has increased the Ca-nadian price of common food productslike milk, poultry, and eggs well abovetheir U.S. counterpart. Tobacco and liquorare both heavily taxed in Canada relativeto the U.S. They are also small in size, highin savings value, and relatively easy tosmuggle by car. Higher prices in the U.S.relative to Canada are then expected todecrease cross border shopping in the U.S.and lower the number of Canadians mak-ing same–day crossings. This will be truewhether the relative price rises for tax orother reasons.

While each commodity class presentsone reason for crossing the border to shop,the individual significance of each com-modity price is not necessary for a test ofthe cross border shopping hypothesis. Theshopping hypothesis relates to the tripand the trip is driven not so much by thesavings made on each individual com-modity as by the aggregate savings real-

ized from the bundle of goods purchased.In this sense, individually adverse pricechanges need not deter border crossingsif the benefits received in the rest of thebundle merit travel. The price predictionthat arises from the model is then that thecoefficient on a price index of some bundleof smuggleable goods [PRBUNDLE] willbe negative. The results given in Table 1are for an index that weighs each com-modity group equally in the aggregate.Experimentation over a wide range of dif-ferent weights, however, produced no sig-nificant difference in findings.16

The exchange rate is often viewed in thepopular press as the key determinant ofcross border shopping and the close cor-respondence of exchange rate movementswith same–day cross border travel in theearly 1990s (as represented in Figure 1)underlies such reasoning. My empiricalwork enters the exchange rate [EXCH] asa separate determinant of cross bordershopping. An increase in the exchangerate, measured in the empirical work asan increase in the number of Canadiandollars needed to purchase one U.S. dol-lar, reduces the incentive to smuggle byincreasing the real cost of U.S. goods toCanadians and so is expected to lead to adecline in cross border travel. The pre-dicted coefficient sign is negative. Thepresence of the exchange rate in a regres-sion that already includes the relativeprice of goods that are most likely to besmuggled allows the regression to iden-tify the separate reasons why a change in

14 Very few Canadians, for example, take the Cornwall crossing back into Canada without filling up at theIndian reservation. The inability to distinguish the origin of the gasoline makes this form of smuggling virtu-ally undetectable. Often smuggling consists of more than simply filling up a full automobile gas tank beforereturning to Canada. The Globe and Mail (August 16, 1993) reports the following tax evasion strategy. A gastanker capable of carrying 40,000 liters of gasoline can pick up two separate half loads of gasoline in the U.S.and declare only half the load coming into Canada. The full load is then sold to Ontario gasoline stations.Then, because gasoline exports from Canada are exempt from the federal tax of 9 cents a liter and the road taxof 15 cents a liter, the returning tanker can purchase gasoline in Ontario and sell it locally while exportingwater to the U.S. The Globe and Mail estimates that this scam can earn $9600 on gasoline never sent to the U.S.

15 As mentioned by one of the referees, the discreteness of mall construction adds a type of hysteresis into crossborder shopping flows, perhaps helping to explain the long adjustment process suggested by the data.

16 Any weighting scheme that increases the size of the weights going to tobacco, gasoline, and liquor productsprices increases the significance of the price of the bundled group. Increasing the weight given to apparel andfood prices decreases the significance of the bundled price.


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TABLE 1CROSS BORDER SHOPPING: ERROR CORRECTION MODEL

WHITE HETEROSKEDASTICITY–CONSISTENT STANDARD ERRORS

Canadian Same–Day Returnings by Automobile from the U.S.: 1972:01–1997:12

D(SAMEDAY)Monthly Differences in same–day

individual crossings

D(AUTOS)Monthly Differences in

same–day vehicle crossings

–269860.2***(1.88)

–554466.5**(2.24)

–798148.7*(3.13)

19546.8***(1.75)

–8969.7(0.122)

89880.3***(1.76)

–1135.2(0.021)

–88340.1***(1.64)

11648.3(0.369)

–0.105*(3.07)

5202449.0*(5.77)

–1293889.0*(5.98)

–1005380.0***(1.75)

31236.9(0.561)

50296.2(1.33)

680797.0*(3.42)

–470374.4**(2.04)

–285543.5(1.16)

508548.4*(2.76)

47525.6(1.53)

–754987.0**(2.29)

–813334.6 **(2.46)

–1288294.0**(2.23)

–2001499.0*(3.40)

48327.0***(1.89)

–125743.1(0.739)

222902.9***(1.90)

13091.9(0.105)

–235350.3***(1.89)

33906.4(0.476)

–0.140*(3.74)

10348586.0*(6.53)

–2250937.0*(5.86)

–2788869.0*(2.94)

85392.8(0.878)

55314.2(0.872)

1352012.0*(4.01)

–1001577.0*(2.61)

–401692.2(1.01)

881362.8*(2.69)

40983.4(0.554)

D(PRBUNDLE)

D(PRBUNDLE(-2))

D(EXCH)

D(EXCH(-1))

D(DEVURATE)

D(HPURATE)

D(USFTA)

D(SUN/ONTARIO)

D(SUNannounced)

D(SUN/QUEBEC)

ERROR CORRECTIONCOEFFICIENT

CONSTANT

PRBUNDLE(-1)

EXCH(-1)

DEVURATE(-1)

HPURATE(-1)

USFTA(-1)

SUN/ONTARIO(-1)

SUNannounced(-1)

SUN/QUEBEC(-1)

February


811

TABLE 1 (continued)CROSS BORDER SHOPPING: ERROR CORRECTION MODEL

WHITE HETEROSKEDASTICITY–CONSISTENT STANDARD ERRORS

Canadian Same–Day Returnings by Automobile from the U.S.: 1972:01–1997:12

D(SAMEDAY)Monthly Differences in same–day

individual crossings

D(AUTOS)Monthly Differences in

same–day vehicle crossings

March

April

May

June

July

August

September

October

November

December

DEPENDENT VARIABLE(–1)







STATISTICS:R2

FQ(14)ProbObservations

227621.7**(2.42)

196982.2**(2.15)

396107.1*(4.56)

211630.7*(2.52)

478604.1*(4.62)

305431.1*(3.45)

–196754.3**(2.36)

–23334.2(0.306)

–119687.5***(1.76)

26952.0(0.413)

–0.316*(5.69)

–0.109**(2.05)

–0.168*(3.08)

–0.204*(3.92)

–0.143*(2.73)

0.253*(5.03)

.91073.959.6.788

299

159730.8*(4.50)

132446.2*(3.65)

184975.4*(4.94)

106504.6*(3.03)

206289.1*(4.814)

123245.5*(3.58)

–8561.1(0.244)

72210.7**(2.24)

29148.7(0.936)

39386.3(1.39)

–0.226*(3.95)

–0.067(1.20)

–0.118**(2.22)

–0.128**(2.19)

–0.140*(2.53)

–0.135*(2.50)

0.283*(5.31)

.86546.5810.7

.710299

t statistics in brackets below the estimated coefficients:* significantly different from zero at 1%, **5%; ***10%

Data Source: Statistics Canada Cansim Database (CANSIM). See Appendix for details.


812

the relative price may induce cross bor-der shopping—i.e., in response to specificprice changes or in response to changesin the relative shopping power of the twocurrencies.17 Because the exchange ratewill change to reflect such factors as in-terest rate differentials and changes in re-source prices, changes in the relative valueof the dollar will capture other relativeprice effects that may be unrelated tochanges in either the tax regime or spe-cialized commodity prices.

The absence of a monthly wage seriesrequires more creative use of availabledata to test for changes in the real valueof time on cross border shopping. Theavailability of the monthly alternative, theunemployment rate, suggests the hypoth-esis that when unemployment is high, therelative lack of market alternatives willlower the cost of time intensive shoppingand hence raise cross border travel. On theother hand, because the unemploymentrate also tracks the business cycle, in-creases in unemployment will reflectlower income and hence times whenshopping has less value. To measure thesize of these offsetting effects I used aHodrick–Prescott filter to separate thechange in the unemployment series intotwo parts: first, the Hodrick–Prescotttrend [HPURATE] taken to reflect incomeeffects associated with permanent levelsof unemployment; and second, the devia-tion of unemployment from its Hodrick–Prescott trend [DEVURATE]. Positivevalues of the latter are then taken to rep-

resent periods of temporary unemploy-ment, where the short run absence of workleaves time available for shopping.18 In-creases in DEVURATE will increase,while increases in HPURATE decrease,cross border shopping.

In addition to these continuous ex-planatory variables, the regressions in-clude a set of discrete variables to accountfor the normal seasonal pattern of bordercrossings between the United States andCanada that may not be related to shop-ping. Weather is the obvious factor under-lying the seasonal pattern of cross bordertravel, explaining the consistent fall insame–day cross border travel each win-ter and the corresponding rise each springand summer. The potential bias that sea-sonal effects might introduce is removedby including centered seasonal dummiesfor each month between February andDecember.19

With this background, we arrive at thediscrete variables used to test the effectspredicted to arise from the regime changesof this time period. They test the premisethat individuals crossing the border totake advantage of relative price differ-ences between countries will also respondto changes in the institutional, regulatory,and tax regimes that underlie the shop-ping environment. Because the free tradeagreement would be expected to lower theadministrative time and inconveniencecost of crossing the border, the analysispredicts that the USFTA should increasecross border travel.20 This is tested by in-

17 Were the exchange rate the only significant price affecting cross border shopping, its inclusion in the regressionwould eliminate the separate significance of the relative price of the shopping bundle [PRBUNDLE]. Whilethis result would not be inconsistent with the cross border shopping story, it would undermine the use of theanalysis to explain crossing flows in response to commodity tax changes that left the exchange rate unaltered.

18 This aligns the shopping hypothesis with theories of labor–leisure choice over the cycle. That is, temporarilylow wages today relative to tomorrow may lead workers to substitute leisure today for more expensive lei-sure tomorrow. Shopping may represent one such use of that time.

19 The centered seasonal dummy subtracts from the traditional monthly (0, 1) value one–twelfth. This centersthe dummy about zero.

20 Note that the dependent variable in the analysis is cross border travel and greater travel need not necessarilycorrespond to greater shopping. Should the free trade agreement have reduced the cost of bringing goods tocustomers legally by more than individuals of evading differential taxes themselves, the presence of USFTAcontrols for agreement induced changes in cross border labor mobility that would otherwise have been attrib-uted to the other shopping variables.


813

cluding in the regression equation adummy variable (USFTA) taking thevalue of 0 for all months through the endof 1988 and 1 from January 1989 onward.A positive coefficient on D(USFTA)would be needed to be consistent with thepredicted permanent rise in the level ofcross border travel following the treaty.

The second major regime change fac-ing Canadian shoppers was the introduc-tion of the GST. In January, 1991 the Ca-nadian federal government removed a13.5 percent manufacturers sales tax andreplaced it with a 7 percent value addedtax on most goods and services.21 WithinCanada, the tax regime change raisedsome final good prices while loweringothers;22 however across the two coun-tries, the presence of the GST raised dis-cretely the final retail price in Canada rela-tive to the price paid for the same good ifpurchased in the U.S. and not declared atthe border when returning. By perma-nently lowering the smuggled relativeprice of U.S. retail goods, the GST is pre-dicted to have permanently increasedcross border shopping by Canadians. Atest for the effect of the GST, however, isalready implicit in the form of the regres-sion equation through the GST inducedchange in the relative price of the bundle[PRBUNDLE]. Should the regime changealso have changed the relative prices ofgoods not in the smuggling bundle andresult in a change in the exchange rate,there would be an additional effect oncross border shopping through EXCH.In the next section, an effort is made toassess this claim and present an estimateof the size of the tax loss associated

arising from cross border efforts to evadethe GST.

Thirdly, the use of the border to evadecommodity taxes such as the GST wouldbe expected to provoke governments sen-sitive to tax revenue losses to respondalong all policy margins, including theinstitutional dimensions of retail marketssubject to different rules and regulations.23

The prohibition of Sunday shoppingwithin Canada in this period, for example,was often given as one reason why U.S.retailers might hold a competitive advan-tage for those Canadian customers livingnear (or adjacent to) the border. Hence,when falling exchange rates and bordercrossing costs led to the major expansionof cross border shopping in the early nine-ties, the devastation of retail trade in Ca-nadian cities close to the border (e.g.,Cornwall and Windsor) placed both localand provincial governments under pres-sure to “level the playing field” by chang-ing traditional retail shopping arrange-ments. The concern over cross bordershopping then became an important rea-son why a number of provincial govern-ments either deregulated aspects of Sun-day shopping (as did Quebec in July, 1990)and/or repealed laws that prohibited latenight and Sunday openings (as didOntario in July, 1993). To test for the sig-nificance of institutional change, the dat-ing of the change poses an interesting ad-ditional test of the shopping hypothesis.For example, in June, 1992, then PremierBob Rae of Ontario, announced his inten-tion to repeal existing Sunday closing leg-islation (implying that existing closinglaws would not be enforced immediately).

21 The major exemptions are food and children’s clothing. See Ruggeri and Bluck (1990) for a more detailedanalysis of the incidence of the GST.

22 Because the GST substituted for another tax (and was designed to be revenue neutral), not all Canadian priceswould rise by the full amount of the tax (i.e., 7 percent). Statistics Canada estimates that 4.4 percent of theprevious average retail store price was accounted for by the federal manufacturers sales tax. On net, the GSTappears to account for a discrete jump in the Canadian consumer price index (CPI) of 3.3 percent (i.e., the CPIrose from 121.0 in December to 125.0 in January, 1991, 1986 = 100).

23 Trandel (1992) explores another dimension of retail competition using spatial analysis. In his case, an openborder and the ability to evade commodity taxation reduces local market power and hence offsets the welfareloss associated with foregone tax revenues.


814

Only in July, 1993 was the new legislationfinally introduced. June, 1992 was thenused as one dummy variable[SUNannounced] to test whether the an-nouncement that Sunday retail store open-ings would not be prosecuted would leadto a decrease in cross border shopping. Ialso used the date (July, 1993) when newlegislation was actually passed [SUN/ONTARIO] as a second dummy variable.Finally, the deregulation of Sunday shop-ping in Quebec was more limited in scopethan Ontario’s, permitting only foodstores with fewer than five (serving) em-ployees to remain open Sundays betweenthe hours of ten and four. To test whetherthat deregulation had a significant nega-tive effect on cross border shopping, atime dummy for the period following July,1990 [SUN/QUEBEC] was included.24

TESTS OF THE MODEL

Table 1 presents the error correction for-mulation of the test of the shopping hy-pothesis. That formulation has the firstdifferences of the explanatory variables(levels for the monthly dummies) and anerror correction term regressed against thefirst differences of two measures of crossborder shopping (same–day individualand automobile crossings). As discussedearlier, the appropriateness of the errorcorrection model is indicated bystationarity in the residuals when the or-dinary least squares (OLS) equation is runin levels.25 The “error” in the error cor-rection model is derived simultaneouslyby nesting the lagged estimating equationin levels within the first difference form(rather than doing the estimation in two

separate steps).26 By using this procedureand presenting the coefficients of both thefirst differences and the error correctionterm, the table allows both the impact andlong run effects of parameter changes tobe discussed. Finally, the equations in-cluded centered monthly seasonal dum-mies and lagged dependent and indepen-dent variables to account for the long andprotracted adjustment process found inthe data. More formally, the stochastic dif-ference equation estimated can be writtenas:

[9]

where yt represents either individual orautomobile daily crossings, xit representsthe set of eight independent variables,and λ represents the error correction co-efficient. Only those lagged variables thatwere found to be significantly differentfrom zero were kept in the equations pre-sented in Table 1.27 While the coefficientsof the contemporaneous first differenceterm measures the short run effect, the co-efficients of lagged variables in levels inthe error correction term need to be ad-justed before the long run effect on crossborder shopping can be isolated. In thiscase the long run effect can be determinedas (–λ)νi.

By inspection, the pattern of results rep-resented by the set of equations as a wholecan be seen to be broadly consistent bothwith each other and with the shoppinghypothesis outlined above. Each equationaccounts for more than 85 percent of the

24 Note that all of the shopping regime dummy variables were given the value 0 in each month before the dategiven in the text and a 1 in the month of change and the months following.

25 The ADF test statistic for the residuals of the SAMEDAY/AUTOS regression equations in levels were – 4.95/–7.42, both of which exceed (in absolute terms) the 1 percent critical MacKinnon value (–3.45). The size ofcoefficient estimate, however, was close to zero, implying an adjustment interval that is long enough to, per-haps, mimic aspects of a unit root.

26 The error correction coefficient estimates when run in two separate stages were –0.090 (2.60) for SAMEDAYand –0.103 (2.44) for AUTOS.

27 The resulting equations minimized the Akaike information criterion.

∆yt = Σβi0∆xi 0 + λ (yt–1 – Σνi xit–1)+ Σβik∆xit–k + Σβj ∆yt–j k = 1, 2

i=8

i=1

i=8

i=8

i=1

i=1 j=1

j=12


815

monthly variation in first differences.Tempering this is the consideration thatmuch of the variability is explained by theseasonal monthly dummies and laggeddependent variables. On the other hand,the inclusion of these terms does result inresiduals that are largely free ofautocorrelation, increasing confidencethat the standard errors are not biased bycorrelation among the residuals.28 Whenone focuses on the set of nine predic-tions—eight variables (plus lags) used inboth the short and long run parts of theequation and the error correction coeffi-cient (all bolded in the table)—fully 30 ofthe model’s 37 coefficient estimates havetheir predicted sign and of these 16 (22)are significantly different from zero whenthe 5 (10) percent level of significance isused.29 More formally, Wald tests of thehypothesis that the set of shopping vari-ables are jointly insignificant can be re-jected at the 1 percent confidence levelfor both the short and long run.30 Finally,the explanatory power of the error correc-tion process is signaled by the signifi-cantly negative error correction coefficientfound in both forms of the test. Thisconfirms the conjecture that a stable sta-tistical relationship exists among thelevels of the variables and hence is con-

sistent with the hypothesis that an equi-librium economic relationship has beenidentified.

Turning to the individual cross bordershopping coefficients in Table 1, both as-pects of the relative shopping price—theunadjusted ratio of bundled commodityprices, PRBUNDLE, and the exchangerate, EXCH—are strongly consistent withthe predictions of the model. In all casesthe coefficients are negative and almostall are significantly different from zero(using the 5 percent confidence interval).31

Comparing the two equations, the num-ber of individuals who cross the borderin response to changes in product pricesvary more than do the number of auto-mobiles, suggesting, perhaps, that it is lesscostly to add an additional shopper to acar than to add an additional car to makeshopping trips. In addition, both sets offindings suggest a long and varied pat-tern of adjustment before equilibrium isreached. The long run coefficients derivedfrom the error correction term, for ex-ample, suggest that the immediate quan-tity response to a relative price change willovershoot the longer run equilibrium. Tosee this, note that the coefficients of theerror correction term must be multipliedby the error correction term to find the

28 The presence of lagged dependent variables in the regression equation means that the Durbin Watson statisticis no longer valid. The alternative Ljung–Box Q–statistic is reported in the table together with the associatedprobability of no serial correlation up to lag length 14.

29 When the equations were run with each commodity price included separately rather than as a tied bundle,some individual price predictions were not supported. As discussed earlier, the model does imply that thebundle should be more significant as a determinant of the shopping trip than any particular commodityseparately. Of the individual commodity coefficients only tobacco and liquor prices were significantly nega-tive as a rule, while the coefficient on apparel was persistently of the wrong sign and often significantlydifferent from zero. The latter is an anomaly in this study and may suggest that the inability to detect thelocation of clothing purchases does not limit effectively the ability to satisfy particular clothing demandswithout increasing the number of cross border trips.

30 The Wald test for the ten (eight) variables describing the short run adjustment process (long run) of theSAMEDAY error correction model yields an F statistic of 4.37 (49.1) and an implied probability of .000003(.0000001) that the variables jointly have no explanatory power.

31 In the exploratory stage of the empirical work I used the current period together with two lags for both thebundled commodity price and the exchange rate to better account for the short run adjustment process. Thelagged variable was then dropped if found to be insignificant. For the bundled price, the first lag was insig-nificant in both forms of the test but was significant at the second lag for individual crossings. In the case ofthe exchange rate, the lagged coefficient was typically larger and more likely to be different from zero than thecontemporaneous value, while the coefficient on the second lag was generally insignificantly different fromzero.


816

long run effect. Doing so results in a longrun coefficient estimate for PRBUNDLEthat is less than half the size of its shortrun counterpart.32 The same kind of re-sult is found for the change in the ex-change rate.33 Together these findings sug-gest that short term border crossings ex-aggerate the mobility results that can beexpected over the longer run. The pres-ence of the second lagged relative priceterm in the individual equation,D(PRBUNDLE(–2)), and the first laggedexchange rate, D(EXCH(–1)) in both, addpersistence to the overshooting responseand suggest that information other thancoordination difficulties will be importantin explaining shopping response throughtime.34

The separation of the unemploymentrate into HPURATE and DEVURATEworks only moderately well as a test ofthe time versus income hypotheses. Tran-sitory deviations from the trend in theunemployment rate, D(DEVURATE), arepositively related to changes in cross bor-der travel as predicted, but both coeffi-cient estimates can be considered signifi-cantly different from zero only under themost liberal interpretation of the confi-dence interval (at 10 percent). Supportingthat more liberal interpretation is the con-sideration that the time hypothesis re-quires the predicted response to be purelytransitory producing additional shoppingonly in the short run. Hence the findingin the two equations that the long run co-efficient on DEVURATE(–1) is consider-ably smaller than the short run coefficientand insignificantly different from zero isconsistent with that hypothesis. Along

these lines it is also interesting to note thatsame–day crossings by individuals aremore responsive to temporary changes inunemployment than are automobile cross-ings. This suggests the severity of the1990–92 Canadian recession as an addi-tional factor in explaining the 1988–93period of widespread cross border shop-ping. Hence, while the number of passen-gers per automobile declined consistentlyacross the entire 1972–98 time period, theepisode of widespread cross border shop-ping by Canadians does stand out as theone time period where this downwardtrend was reversed, even if only tempo-rarily. Changes in the Hodrick–Prescotttrend, however, are not consistent with thehypothesis that permanently higher lev-els of unemployment reduce householdincome and so decrease cross border shop-ping. In both equations the short run co-efficient, D(HPURATE), has the expectednegative sign, but neither the short nor thelong run coefficients are significantly dif-ferent from zero.35

It is in relation to the tests of institu-tional and shopping regimes that theequations present their most interestingpositive findings. First, the data is stronglyconsistent with the hypothesis that theU.S./Canada Free trade agreement,USFTA, by reducing the time and otherbureaucratic costs of crossing the border,has increased same–day cross bordertravel and the potential for cross bordershopping. This is reflected somewhat inthe size and significance of the short runadjustment coefficients, but is representedeven more strongly in the persistence andsize of the long run coefficients of the er-

32 The long run coefficient associated with PRBUNDLE(–1) is –315131.2 which is less than half of that estimatedfor ∆PRBUNDLE (–754987.0).

33 Here the long run coefficient estimate is –390,442.6, which is roughly one–third of the short run adjustmentcoefficient of –1288294.6.

34 Recent work by Vilasuso and Menz (1998) suggests that individuals respond asymmetrically to changes indomestic versus foreign variables. Their analysis highlights information costs in predicting the size andtiming of cross border flows.

35 It does remain true, however, that separation does provide a better fit with the data than does the singleaggregate and to the extent that this division does allow a better test time and income hypotheses, the data isconsistent with the time hypothesis but not the income hypothesis.


817

ror correction term.36 The U.S./CanadaFree Trade Agreement, soon followed bythe North American Free Trade Agree-ment, resulted in a large increase in same–day cross border travel. What is less cer-tain is whether that larger flow actuallyreflects an increase in cross border shop-ping.37

The Sunday shopping hypothesis offersan alternative test of the time cost of shop-ping and here the error correction formu-lation captures some of the most interest-ing features of the data. Of the threedummy variables, SUNannounced, theannouncement of the political decision notto enforce Sunday closing laws in Ontario(followed by widespread retail store open-ing) might have been expected to havehad the biggest effect on shopping con-venience in the short run and hence tohave produced the largest fall in the costof shopping domestically relative to cross-ing the border.38 As expected, the datadoes suggest that shoppers did react im-mediately to the decision not to enforcethe Sunday restrictions rather than waituntil their formal removal a year later,SUN/ONTARIO. The short run an-nouncement coefficient, D(SUN/an-

nounced), dominates in size and signifi-cance the date at which the legislation wasfinally passed D(SUN/ONTARIO). Whenthe long run coefficients in the error cor-rection term are examined, however, thetwo are reversed and the date of legisla-tion can be seen to have greater explana-tory power.39 In part this may be becausemany Ontario retailers would commitfully to longer term reorganization onlywhen the opportunities presented by latenight and Sunday shopping were recog-nized to be permanent.40 The generaltenor of the findings reinforces earlierobservations on the importance of wheninformation arrives and adds to this thedistinction between temporary and per-manent change for explaining shoppingbehavior and hence the timing of crossborder flows. Finally, neither equationshowed any significant cross border re-sponse to the deregulation of mid– tosmall sized food stores in Quebec in theshort run, D(SUN/QUEBEC). Both shortrun coefficients are insignificantly differ-ent from zero. In the long run, however,the results are perverse—cross bordertravel increases despite increasing domes-tic shopping convenience.41

36 The long run coefficient of 189,281 is statistically indistinguishable from the short run coefficient of 222,903.This is one exception to the general rule that short run effects are larger than long run effects.

37 In an earlier version of the paper I noted that USFTA produced a stronger empirical effect on same–daycrossings using the month before the treaty was actually implemented. Because new regulations would notgenerally be implemented before the treaty actually came into effect, this would suggest either that shoppingwas not the only cross border activity enhanced by the prospect of a treaty or that cross border shopping relieson information and attitudes towards cross border travelers that depend as much on attitude as they do onthe enforcement of formal rules and regulations.

38 Ontario represents about 40 percent of Canada in terms of both population and income while Quebec is about25 percent. In addition, Quebec’s deregulation in 1990 covered a relatively small subset of retail stores (middleto small grocery stores).

39 While all four long run coefficient estimates have the expected negative sign, the coefficients of the date oflegislation (SUN/ONTARIO) are double the size of the announcement dummy and both significantly differ-ent from zero (using a 5 percent confidence interval). Comparing the short and long run, the short run an-nouncement coefficient is roughly a third larger than the long run effect of legislated Sunday closings inOntario (140,221).

40 The year between the Rae announcement and the passing of new store hour legislation was one of intensedebate over the form and substance of the proposed legislation. In general labor unions argued againstgreater liberalization, communities adjacent to the border argued for Sunday shopping, and large retailerssuch as Eatons and the Bay split on the issue. See Ferris (1991).

41 This is the only shopping hypothesis directly contradicted by the data. Combined with other anomalousfindings (see footnote 8), a more detailed study of Quebec’s cross border travel seems warranted and wouldmake an interesting special study.


818

Aside from the case of Quebec, the tim-ing of regulatory change with respect toshopping in Canada and its effect on crossborder travel suggests the deliberate useof regulation by some Canadian govern-ments to stem the flood of cross bordershopping that followed the dramatic risein the value of the Canadian dollar rela-tive to the U.S. dollar and peaked with theadoption of the GST. Without suggestingequivalency, these finding are consistentwith the hypothesis that institutional/regulatory change complement and some-times substitute for more traditionalpolicy incentives to guide individualbehaviour (such as through tax reductionsor greater enforcement activity). Viewedfrom a broader perspective, these findingssuggest that greater Canada/U.S. open-ness may have begun a competitionamong political jurisdictions that has re-sulted in greater uniformity in a widerrange of shopping characteristics. Notonly may open borders bring tax rates andstructures into greater conformity, but po-litical competition may also induce uni-formity in organizational structures andregulatory practices. With either closedborders or costly crossings, countries havegreater scope to bundle the package ofservices, taxes, institutions and regula-tions that best suit their average constitu-ent.42 Open borders, on the other hand,induce competition not only at the levelof the bundled aggregate but also alongeach of the bundle’s margins. Becausecompetition among jurisdictions is alwaysfor the marginal rather than the averageindividual, open borders privilege thetastes of the most mobile. From this per-spective it is not always clear that greatermobility will enhance the efficiency ofgovernmental institutions.

Finally, the error correction coefficientsin each equation are negative, as pre-dicted, and both are significantly differ-

ent from zero. As an indicator of the rateof adjustment, the small size of the coeffi-cient estimates suggests that adjustmentto long run equilibrium will be long andprotracted. The protracted nature of ad-justment is also indicated by the numberand pattern of lagged dependent variableterms that were found to be statisticallysignificant in the short run part of themodel, reinforcing the earlier finding thatshort run coefficients were typically largerthan their long run counterparts. Togetherthese findings underline a theme that hasarisen at various earlier stages in theanalysis—that is, the need to better un-derstand the role of information and othertransactions costs to explain the lengthand pattern of the adjustment process.

ECONOMIC SIGNIFICANCE

In the previous section I test for the pre-dicted sign and statistical significance ofthe key shopping variables and find re-sults that are strongly consistent with themodel. In this section I turn to the ques-tion of economic significance by measur-ing the size of the estimated effects oncross border travel following substantivecommodity tax changes and ask the re-lated question of whether the magnitudesinvolved imply a tax revenue loss substan-tial enough to merit explicit policy atten-tion. To focus this analysis I use the firstequation of the error correction model inTable 1 to assess the size of the impact andlonger run effect of the GST.

The quantitative response to the GST isimplied by the estimated response by in-dividuals to actual relative price and othervariable changes that followed the impo-sition of the GST in January of 1991. Allstudies of the impact of the GST find thatthe substitution of the GST for the Fed-eral Manufacturer’s sales tax did result inan overall rise in consumer prices and that

42 This conclusion is particular to the theory used to explain the size of government in the model. Note that ifgovernment were a Leviathan, then competition constraining its size would confer benefits rather than costs.


819

this rise was spread over the first fewmonths of 1991.43 Using the short runPRBUNDLE coefficients and attributingall of the actual change in PRBUNDLE tothe GST, the model implies that the GSTresulted in an increase in cross bordershoppers of slightly less than 100,000 permonth or about 1,140,000 per year.44 Thesubstitution of a broad 7 percent valueadded tax for a thirteen percent man-ufacturer’s sales tax would have also beenexpected to lead to an increase (decrease)in Canadian exports (imports), strength-ening the value of the Canadian dollar andlowering the exchange rate. On the em-pirical side, however, few exchange ratestudies have attributed any change in theexchange rate to the GST, rather attribut-ing the rise in the value of the Canadiandollar in that period to restrictive Bank ofCanada monetary policy.45 If I attribute allof the January/February 1991 fall in theexchange rate to the GST, the equationcoefficients would imply an additional160,000 crossings per month.46 Using theabsence of any exchange rate effect as alower bound and its total inclusion as anupper bound, the estimates of the errorcorrection model suggest a short run in-crease in the number of cross border shop-pers of between 100 and 250,000 permonth (or 1–3 million additional crossingsper year). As discussed above, the modelsuggests that the long run effect will bemuch smaller. Treating the change in rela-tive prices and exchange rates by March,1991 as the appropriate measure of the

permanent change following the GST, themodel suggests at most 50,000 additionalcross border shoppers per month (600,000per year).47

While the absolute size of the travel re-sponse in the short run underlines thequantitative importance of our earlier sta-tistical findings, the total tax revenue lossthat follows a new tax program is muchharder to estimate. In terms of tax lossesarising from cross border shopping, it in-cludes more than just the losses arisingfrom consumers who now choose to sub-stitute cross border shopping for domes-tic retail sales. Rather, the total cross bor-der revenue loss will include theintramarginal foregone sales tax from thepurchases of individuals who alreadycrossed the border to shop. However, onlyfor sales tax programs already in existence(such as for existing provincial sales taxes)will the new tax losses arising from addedcross border shopping be due entirely tothe GST.48 Together with the full tax rev-enue loss arising from the marginal shop-per, and even though the ability to avoidthe GST was not necessary as an incen-tive for intramarginal cross border shop-ping, their unwillingness to voluntarilydeclare cross border purchases on theirreturn will further expand the loss in taxrevenues for the federal government.

Estimates of personal expenditures bysame–day Canadian travelers in theUnited States are made quarterly by Sta-tistics Canada from voluntary responsesto questionnaires handed out to a sample

43 The Bank of Canada (1991), for example, estimates that the GST accounted for a 2.22 percent rise in the CPI byMay of 1991. The delay in the full impact of the GST is due, in part, to many retailers choosing to incorporate“we’ll pay the GST” sales into the post–Christmas retail sale period. See also Clancy and Smith (1991).

44 By March of 1991, the short run effect of the actual change in PRBUNDLE on cross border travel was(–753987)(–.098) + (–813335)(–.026) = 73891 + 21147 = 95038 individual crossings per month or 1,140,456 per year.

45 See, for example, LaFrance and van Norden (1995).46 That is, (–.045)*(–1288294) + (–.055)*(2001499) = 160,923.47 In this case (–.055)*(–390441) + (.098)*(315131) = 52357.48 Canadian border crossing stations did not initially collect provincial sales taxes. With the Liberal government’s

election after the GST’s introduction, however, the federal government negotiated tax harmonization agree-ments with Quebec (in 1991 and 1992) and the Maritime provinces (1997) to coordinate sales tax collectionsinternally and at the border. Only in March, 1999 did the Federal government start to collect Ontario’s salestax at the border.


820

of returning travelers.49 Because individu-als cannot be expected to self report evi-dence of illegal activities (despite prom-ised anonymity), such data can be ex-pected to contain a serious under report-ing bias. Nevertheless, using that estimateof average same–day expenditures of$29.21 per person, the 7 percent GST lossper trip would be roughly $2.00.50 On thisbasis, the short run incremental tax lossdue to induced cross border shoppingwould have been between 2 and 500,000a month (or $2.5 to 6 million dollars ayear). Since these new cross border shop-pers will not pay provincial sales taxes, aroughly equivalent tax loss will arise forthe provinces.51 The intramarginal GSTloss is much larger. With slightly morethan $59 million same–day border cross-ings in 1991, the incremental GST tax losswould have been in the order of $110 mil-lion per year. Together these figures sug-gest a GST tax loss of $120 million dollarsa year.52

It does seems likely, however, that theability to realize a $2.00 tax savings bycrossing the border to shop would be in-sufficient to overcome inertia let alonecover the inconvenience and other costsof cross border shopping. Such savingswould not generate the significant priceresponse implied by the equation esti-mates.53 Put in a slightly different way, asingle gasoline fill up of 50 imperial litreswould have led to a $5.00 saving to eachCanadian shopper and, even without

counting the GST levied on gasoline sales,tax revenue losses to federal and provin-cial governments in Canada of over$10.00.54 Similarly, the tax savings (loss)made by the consumer (Canadian govern-ment) when bringing back even one car-ton of (200) cigarettes was over $25. It isthen not unreasonable to expect a highertax revenue loss than the self reported fig-ure of $2.00 used above. On the basis of$10.00/person, the aggregate tax loss im-plied by this analysis would be $600 mil-lion per year in the short run. This corre-sponds to about 4 percent of the $15.2 bil-lion in GST collections realized in the1991/92 fiscal year.55

While the tax loss associated with indi-viduals not declaring U.S. goods whencrossing back into Canada is large, theconfluence of a transitory change in theexchange rate, particularly heavy exciseduties on particular products (such as to-bacco) and the new federal GST also ledto an extraordinary period of organizedsmuggling and further tax loss. It is un-clear exactly how much of the smugglingeffect can be attributed directly to the GST;nevertheless, it is clear that the unpopu-larity of the GST served to legitimizesmuggling flows and that the size of thetax losses due to smuggling were substan-tial. Direct evidence on the scale of (pro-fessional) cross border tax evasion comesfrom the Canadian Auditor General’s 1996special review of the Excise Tax Act in re-lation to four specific products in the early

49 See Kemp (1992) for a description of the survey methods and conceptual issues involved in measuring theseflows.

50 Kemp (1992), Table 1, Column (2), p.5.7.51 Except for Alberta, all provinces have sales taxes running between 6 and 12 percent (as of August, 1991).

Ontario’s provincial sales tax rate was 8 percent, implying a cross border tax loss slightly larger than thefederal GST tax loss.

52 The permanent loss associated with each case is roughly one–fifth of the short run size.53 In 1991 the large volume of cross border traffic often led to two to three hour waits on the weekends to pass

through customs.54 In August 1991, the federal excise tax was 8.5 cents per litre and provincial taxes ranged from 13.7 and 9 cents

a litre. The further addition of the GST (and often provincial sales taxes) to the excise tax inclusive pricemeans that roughly one–half of the pump price goes to taxes. See Bank of Canada, p.16. The difference forconsumer savings comes from the presence of U.S. taxes.

55 The Budget 1992, February 25, 1992, Department of Finance, Ottawa p. 104 and Statistics Canada, Cansim(Statistics Canada Database) series number D93729, Excise Taxes and Duties: Goods and Services Tax.


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nineties: tobacco, alcohol, jewelry, andmotive fuels.56 In that report the AuditorGeneral finds that due to smuggling, to-tal tobacco tax revenues actually declinedbetween 1991 and 1994, with tax lossespeaking at over $2 billion in 1993 (18–10).The Auditor General also reports that“The major problem with respect to alco-hol is the evasion of excise duty . . .accomplished largely through smuggling. . . and by the diversion of exportsand illegal production” (18–10). The bor-der also played a significant role for mo-tor fuels as evasion arose “through the di-version . . . from tax–exempt to taxableuses, . . . of products destined for export,and illegal production” with estimatedlosses of roughly $55 million in 1993 (18–10).

In some respects the 1991–93 periodwas anomalous, particularly for tobaccosmuggling. However, even following thedramatic cutting of federal and provincialtax rates in February 1994, the tax revenuelosses due to cross border smuggling re-main a significant concern. Table 2 pre-sents the Auditor General’s estimates ofrevenue loss from tax evasion for eachproduct relative to tax collections. Asmuch as 10 percent of potential revenuecontinues to be lost due to tax evasion,

with a substantial part of that loss com-ing from cross border smuggling.57 In ag-gregate terms, the tax loss arising fromorganized smuggling activity is at leastthree times as large as the losses arisingfrom individual’s cross border shopping.

While it is tempting to conclude fromthis exercise that cross border shoppingin its most general sense is a significantconstraint on tax policy for Canada, theproblematic nature of many of our as-sumptions calls for somewhat greater cau-tion. What does seem unambiguous, how-ever, is that the combination of individualcross border shopping and professionalsmuggling produced quantitatively sig-nificant tax losses during the implemen-tation period of the GST. As important, thesignificance of the response of cross bor-der travel to the prices of goods that canbe smuggled questions the size and im-portance of self–reported average expen-diture data (for same–day cross bordertravel) reported by Statistics Canada andused in policy analysis (e.g., Di Matteo,1993; and Boisvert and Thirsk, 1994). Theexpectation that in voluntary surveys in-dividuals will have “no valid reason tomisreport, since the form is completelyanonymous”(Kemp, 1992, p. 5.5) does notseem justifiable for activities that are ille-

TABLE 2TAX REVENUES, ESTIMATES OF REVENUE LOSS FROM EVASION,

AND EVASION AS A PERCENTAGE OF TAX COLLECTIONS: 1994/95

CommodityTax Revenues($ millions)

Estimate of Tax Loss($ millions)

Revenue Loss as %of tax collections

Motive FuelsTobaccoSpiritsTotal

3,819.51,918.1

366.16,103.7

55 to 110200 to 280150 to 200405 to 590

1 to 310 to 1440 to 557 to 10

Source: Report of the Auditor General of Canada to the House of Commons, Chapter 18, Revenue Canada andthe Department of Finance—Excise Duties and Taxes on Selected Commodities, Exhibit 18.3 (18–11), Ottawa,1996.

56 Report of the Auditor General of Canada to the House of Commons, Chapter 18, September, 1996. Note thatthe Auditor General evaluates different aspects of tax structure each year so that a time series of estimates oftax losses through evasion is not possible. This report covers some sources of GST tax evasion for the years1991 through 1994.

57 Note how the estimates of percentage tax duty losses by commodity reinforce our findings for cross bordertravel. That is, given this table it is not surprising to find same–day cross border travel responding strongly tochanges in tobacco and alcohol prices.


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gal. Individuals should be expected to un-derstate their involvement in these activi-ties, partly because they expect to repeatthem in the future.

CONCLUSION

What, then, has accounted for the dra-matic rise and fall of cross border shop-ping that took place in Canada between1989 and 1994? This analysis suggestsmany causes: the substantive fall in theCanadian exchange rate (that proved tobe transitory); a particularly sharp rise inunemployment rates associated with the1990–92 recession; increasingly heavytaxes on particular “sin” products inCanada relative to the U.S. (such as liquorand tobacco); and the arrival of a free tradeagreement that accelerated the adoptionof a value added tax, discretely increas-ing the price of final retail goods inCanada relative to the U.S. The size of thetax loss produced by cross border tax eva-sion, discussed in the previous section,provides at least one reason for the dra-matic decline in cross border shopping.That is, the size of the tax losses and smug-gling activities associated with tobacco ledCanadian governments to dramaticallylower tobacco taxes (halving tobaccoprices). This had an immediate impact onboth professional tobacco smuggling andindividual cross border shopping.58 Assignificant has been the rise of the ex-change rate back to (and beyond) its pre–1988/89 level. Even with these changes,however, the continued importance of theGST as a revenue source suggests thatcross border shopping will continue to

constrain the ability of the federal andprovincial governments to use sales taxesto tax Canadians differentially.

Even though the major portion of therevenue loss from cross border shoppingarises from professional smuggling,changes in cross border travel may regis-ter changes in the intensity of cross bor-der tax evasion on the political processmore effectively than do foregone taxes.The demonstration effect of hundreds ofthousands of Canadians clogging bordercrossing stations seems to have been suc-cessful in politically capping the GST andmay have been instrumental in inducingregulatory change and changing taxpolicy. The liberalization of Canada’sshopping hour regulations has alreadybeen noted, but even the composition oftax revenue collections may have beenchanged by this experience. Prior to theperiod of widespread cross border shop-ping, provincial sales taxes and a varietyof “sin” taxes were popular fundingchoices for new government initiatives.Since the Free Trade Agreement and theadoption of the GST, payroll taxes havecome to replace sales taxes as the tax in-strument of choice.59 This is consistentwith the belief that payroll taxes offerfewer opportunities for evasion.60

All of this suggests that ever greatercross border mobility between the U.S.and Canada has increased the cost of sus-taining significant differences in corpo-rate, sales, and income taxes. Whetherthese considerations help to explain therecent pressure on the size and rate ofgrowth of government requires furtheranalysis. However, it seems safe to con-

58 Again using the parameters of the model, the rise in PRBUNDLE from 1.46 to 1.72 in February, 1994 resultedin a short run fall (over three months) in the number of cross border trips of slightly over 400,000 per monthand a long run response about 80,000 fewer trips per month.

59 The increasing importance of payroll taxes as a revenue source for governments was the subject of a forum atthe 1998 Canadian Economic Association Meetings in Ottawa, since published in the September, 1998 issue ofCanadian Public Policy. The William Robson and William Scarth introduction setting the tone for that sessionbegins, ”[T]he upward trend in Canadian payroll taxes over the years, and the prospect of further hikes, raisesa number of questions.”

60 A recent concern within policy circles in Canada over brain drain losses to the U.S. suggests that payroll andincome tax differences are increasingly recognized as allocative (cross border) in their effect.


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clude that open borders have increasedthe cost of maintaining a rate of growthfor government in Canada that exceedsthe rate of growth in the U.S. Whether thisprediction for government size translatesinto a gain or loss in Canadian welfare ismore difficult to assess. The constraint onpublic goods implied by the ability toevade taxation using the border restrictsCanadian welfare if the level of provisionchosen by the median voter mirrors thechoice of the average Canadian. On theother hand, welfare will rise if the medianvoter bears less than his or her share ofcost and imposes too high a level of pub-lic output.

Acknowledgments

I would like to thank Keith Acheson, JeffBernstein, Alok Bohara, Brian Erard, Soo–Bin Park, Eddie West, and particularlyJohn McManus for volunteering their ex-pertise on specialized aspects of cross bor-der smuggling. I am also indebted to threereferees from this Journal for suggestingimprovements in both content and style.Finally, I would like to acknowledge thehospitality of the Department of Econom-ics at the University of New Mexico, wherethis version of the paper was first written.

REFERENCES

Bank of Canada.“Targets for Reducing Inflation: FurtherOperational and Measurement Consider-ations.” Bank of Canada Review (September,1991): 3–26.

Boisvert, Michelle, and Wayne Thirsk.“Border Taxes, Cross–Border Shopping, andthe Incidence of the GST.” Canadian Tax Jour-nal 42 No. 6 (December, 1994): 1276–93.

Clancy, D., and P. Smith.“The Allocation of Indirect Taxes and Sub-sidies to Components of Final Expenditure.”National Income and Expenditure Accounts 39No. 1. Statistics Canada Catalogue 13-001,1991: xxxiii–liii.

Di Matteo, Livio.“Cross–border Trips and Spending by Ca-nadians in the United States: 1971–1991.”Canadian Business Economics 3 No. 4 (Octo-ber, 1993): 51–61.

Ferris, J. Stephen.“On the Economics of Regulated Early Clos-ing Hours: Some Evidence from Canada.”Applied Economics 23 No. 8 (August, 1991):1393–400.

The Globe and Mail.(August 16, 1993): A7.

Kanbur, Ravi, and Michael Keen.“Jeux Sans Frontières: Tax Competition andTax Coordination When Countries Differ inSize.” American Economic Review 83 No. 4(September, 1993): 877–93.

Kemp, Katharine.“Cross–Border Shopping—Trends andMeasurement Issues.” Canadian EconomicObserver 11–010 (December, 1992): 5.1–5.13.

Lafrance, Robert, and Simon Van Norden.“Exchange Rate Fundamentals and the Ca-nadian Dollar.” Bank of Canada Review(Spring, 1995): 17–32.

Lovely, Mary E.“Crossing the Border: Does CommodityTax Evasion Reduce Welfare and CanEnforcement Improve It?” Canadian Jour-nal of Economics 27 No. 1 (February, 1994):157–74.

Lovely, Mary E.“Optimal Commodity Taxation with CostlyNoncompliance.” Public Finance Quarterly 23No. 1 (January, 1995): 115–30.

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Robson, William, and William Scarth.“The W. Irwin Gillespie Round Table onPublic Policy II. Payroll Taxes: Pros andCons; Ups and Downs.” Canadian PublicPolicy 24 No. 3 (September, 1998): 379–401.

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Thursby, Marie, Richard Jensen, and JerryThursby.

“Smuggling, Camouflaging, and MarketStructure.” Quarterly Journal of Economics 106No. 3 (August, 1991): 789–814.

Trandel, Gregory. “Evading the Use Tax onCross–Border Sales: Pricing and Welfare Ef-fects.” Journal of Public Economics 49 No. 3(December, 1992): 313–31.

Appendix

Sources of Canadian Monthly Data: 1972.01–1997.12

The data in the empirical tests comes primarily from CANSIM: the Statistics Canada Database.The CANSIM series numbers are presented below.

1. Border Crossing Data:D145037 Canadian residents returning by auto, same-day SAMEDAYD146225 Canadian automobiles returning, same-day AUTOS

2. Price Data:B3400 Exchange rate measured as # of Canadian $ per U.S. $ EXCH or EXCHANGERATE

P100000 CPI Canada (1992 = 100), All itemsP100002 CPI Canada, Price of Food purchased in storesP100363 CPI Canada, Gasoline PriceP100275 CPI Canada, Clothing PriceP100266 CPI Canada, Tobacco PriceP100257 CPI Canada, Alcoholic Beverage PriceD139105 CPI U.S. (1982–4 = 100), All itemsD139107 CPI U.S., Food PriceD139120 CPI U.S., Gasoline PriceD139118 CPI U.S., Apparel PriceD139125 CPI U.S., Tobacco PriceD139109 CPI U.S., Alcoholic Beverage Price

3. Relative PricesPRAPPAREL = [D139118/P484275]PRFOOD = [D139107/P484002]PRGAS = [D139120/P484363]PRTOBAC = [D139125/P484475]PRLIQUOR = [D139109/P484478]PRBUNDLE = .2*PRAPPAREL + .2*PRFOOD + .2*PRGAS + .2*PRTOBAC + .2*PRLIQUOR

4. Other Cansim DataD767289 and D980404 Canadian Unemployment Rate (Both sexes 15+) URATE

5. Dummy VariablesUSFTA 0 through December 1988, 1 thereafterSUNannounced 0 through May 1992, 1 thereafterSUN/ONTARIO 0 through June 1993, 1 thereafterSUN/QUEBEC 0 through July 1990, 1 thereafter

Trandel, Gregory.“Interstate Commodity Differentials and theDistribution of Residents.” Journal of PublicEconomics 53 No. 3 (March, 1994): 435–57.

Vilasuso, Jon and Fredric Menz.“Domestic Price, (Expected) Foreign Price,and Travel by Canadians in the UnitedStates.” Canadian Journal of Economics 31 No.5 (November, 1998): 1139–53.

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