By estimating a staggered price model over the period 1980q1-2010q2, this paper documents that, after the euro changeover, Italian retailers have increased the number of price adjustments, which has translated into a higher inflation rate, with a detrimental effect on the competitiveness of the Italian economyEuro changeover, staggered price adjustments, inflation

Alessandro Girardi

Guglielmo Maria Caporale

Marco Ventura

English

Research Papers in Economics

 
 
 
 
 
 
 
 
The Euro Changeover and Price Adjustments 
in Italy 
 
 
 
Guglielmo Maria Caporale 
Alessandro Girardi 
Marco Ventura 
 
 
CESIFO WORKING PAPER NO. 3386 
CATEGORY 7: MONETARY POLICY AND INTERNATIONAL FINANCE 
MARCH 2011 
 
 
 
 
An electronic version of the paper may be downloaded  
• from the SSRN website:              www.SSRN.com 
• from the RePEc website:              www.RePEc.org 
• from the CESifo website:            Twww.CESifo-group.org/wpT CESifo Working Paper No. 3386 
 
 
 
The Euro Changeover and Price Adjustments 
in Italy 
 
 
Abstract 
 
By estimating a staggered price model over the period 1980q1-2010q2, this paper documents 
that, after the euro changeover, Italian retailers have increased the number of price 
adjustments, which has translated into a higher inflation rate, with a detrimental effect on the 
competitiveness of the Italian economy. 
JEL-Code: C220, E310. 
Keywords: Euro changeover, staggered price adjustments, inflation. 
 
 
 
 
 
  
Guglielmo Maria Caporale 
Centre for Empirical Finance 
Brunel University 
UK – West London UB8 3PH 
Guglielmo-Maria.Caporale@brunel.ac.uk 
  
Alessandro Girardi 
Ministry of Economic Affairs and Finance 
Rome / Italy 
a.girardi@isae.it 
Marco Ventura 
Italian National Institute of Statistics 
Rome / Italy 
m.ventura@isae.it 
 
 
March 2011 2 
1. Introduction 
On the 1
st of January 1999, national currencies were replaced by the euro in eleven European 
countries. According to some economists, the switch to euro-denominated prices for goods and 
services (the euro changeover) did not amount to a mere change of the numeraire for current 
transactions: it could also have generated speculative behaviour, through the implicit coordination 
of price setters towards higher price equilibria, as a result of money illusion or market imperfections 
(Fehr and Tyran, 2001; 2007). 
Although the possible effects of the euro changeover have been debated in all the countries that 
switched to euro denominated prices, this has been a particularly controversial issue in Italy, where 
it has involved consumers claiming that official statistics were not reporting the “true” inflation 
rate, the authorities and academics as well. Marini et al. (2007), among others, have argued that the 
combined effect of the introduction of the new currency and the existence of industries with market 
power could have produced self-fulfilling inflationary expectations leading to discontinuous price 
jumps totally unrelated to underlying market conditions or fundamentals. 
A possible explanation for the difference between perceived and actual inflation is that consumers 
attach greater weight to price changes in goods and services bought more frequently relative to the 
so-called ‘big ticket items’, such as durable goods (ECB, 2003; Marini et al., 2007). Official 
inflation measures reflect instead the price changes faced by hypothetical average consumers, 
whose consumption basket matches the consumption structure of the economy as a whole. The 
present paper aims to provide some empirical evidence on whether there exists a changeover effect 
in the official inflation measure in the specific case of Italy.  
In our view, quantifying the possible impact of the changeover effect amounts to identifying a 
discontinuity in sellers’ pricing behaviour in the changeover period: since price adjustments are 
costly (Zbaracki et al. 2004; Bergen et al., 2008), a greater number of adjustments should imply an 
acceleration in inflation dynamics. For this purpose, we estimate a pricing function based on a 3 
staggered pricing model that has already been used successfully for the US (Galì and Gertler, 1999, 
Hall et al. 2000; Bakhsi et al 2007; Janko 2008). Our results document an increase in the average 
number of price changes, the euro changeover effect being found equal to around 40% of the 
inflation rate.  
2. Empirical Analysis 
2.1. Theoretical underpinnings 
While in the rational expectations paradigm prices are perfectly flexible, in staggered pricing 
models (Fisher, 1977; Taylor, 1979; Calvo, 1983) they are adjusted by price setters towards 
optimising levels at discrete time intervals. Indeed, in the real world continuous re-optimising of 
price levels is not a viable option for firms, since it entails costs related to information gathering, 
evaluation, price decision and revision of the price list (Akerlof and Yellen, 1985; Akerlof, 2002).  
A common feature of staggered price models is the inclusion of a constraint on the frequency of the 
price optimisation process, with the overall inflation rate being seen as the result of aggregating 
individual price setting decisions. This aggregation is greatly simplified by Calvo’s approach 
(1983), which assumes that in a given period there is a fixed probability, θ, that a firm will 
maintain its price constant. Therefore, in any given period the fixed probability that a price setter 
will adjust his price in the same period is  − θ 1 . Following the approach of Galì and Gertler (1999) 
and Eichenbaum and Fisher (2003), we estimate the following equation: 
{[ ( )( ) ] } tt t tt Es z
−
+ π− θ − θ − β θ − β π =
1
1 11 0         ( 1 )  
where  t E is the expectation operator at time t ,  t π  and  t s  represent (the deviation from the steady-
state of) the inflation rate and real marginal costs, respectively, β is the subjective discount rate, 
and  t z  is a set of instruments that must satisfy the orthogonality condition. 
Testing whether inflation accelerated after the euro changeover implies testing that a smaller 
fraction of price setters kept their price unchanged, i.e. that θ  was lower on average. In turn, this 4 
means a shorter average period during which prices are unchanged, as given by  ()
− ξ≡ −θ
1 1 . In 
order to capture possible changeover effects, we estimate equation (1) including a dummy variable, 
t D, taking value 0 for the quarters between 1999q1 and 2003q4, and 1 otherwise, and then we test 
the constancy of θ  over the different periods.
1 The empirical model is therefore specified as 
follows: 
{[ ( ) ( ) ( ) ( ) ( ) ] } tt t t tc c c t tt ED sD s z
−−
+ π− ⋅ θ ⋅ − θ⋅ − β θ⋅ − − ⋅ θ ⋅ − θ ⋅ − β θ ⋅ − β π =
11
1 11 1 1 1 0  (2) 
2.2. Data and estimation results 
The data are taken from the OECD Main Economic Indicators database. The sample consists of 
quarterly observations over the period 1980q1–2010q2. We consider four alternative aggregate 
price series: 1) the Consumer Price Index (CPI), 2) the consumption expenditure deflator (CED), 3) 
the GDP deflator (GDPD), and 4) the Harmonized CPI (HCPI). Real marginal costs are computed 
as the logarithm of the wage share of GDP. Following Galì and Gertler (1999) and Eichenbaum and 
Fisher (2003), we use the sample period average for the four measures of inflation and for real 
marginal costs as a proxy for their respective steady-state values. As for the set of instruments, we 
include a constant term, labour income share, quadratically detrended real GDP, the spread between 
the annual interest rate on the ten-year Treasury bond and three-month Treasury bills, and the 
growth rate of the producer price index and of nominal wages. This corresponds to the basic set of 
instruments used in Gali and Gertler (1999). 
GMM estimation results for equation (2) are presented in Table 1. A HAC Newey-West estimation 
weighting matrix was used (Greene, 2008). Given the difficulty of estimating β with precision 
                                                            
1 Riaño et al. (2007), among others, identify the euro changeover in a similar way. The chosen ending period (namely, 
2003q4) includes some delayed effect in sellers’ price revisions. Similar results are obtained with the 1999q1-2004q4 
window.  5 
across the different inflation measures, we follow Eichenbaum and Fisher (2003) by setting this 
parameter equal to 0.99. 
[Table 1] 
The comparison between θ  and  c θ shows a sharp decrease in the share of firms keeping their retail 
prices fixed after the changeover, and this holds for each inflation measure. In the changeover 
period price changes occur after 1.1 quarters, a much lower value than in the previous period (i.e., 
1.9 - 1.8 in the case of GDPD). The change in the value of θ implies an average increase in the 
number of adjustments per year, given by  () c
−− λ ≡⋅ ξ− ξ
11 4 , ranging from 1.4 to 1.7. Finally, the 
Wald test rejects the null of no break for all the inflation measures whilst the Sargan J statistics 
validates the set of instruments. 
2.3. Assessing the changeover effect 
In order to translate the extra number of adjustments made (λ) into a quantitative estimate of the 
euro impact on actual inflation, we recall that price revisions entail “small menu costs” for firms. 
According to Levy et al. (1997), such costs amount to around 0.7% of total annual revenues. 
Assuming that the cost of price adjustments represents at least this percentage of annual revenues, 
then during the changeover period total revenues should increase by at least 0.7 times λ for sellers 
to break even. 
Total revenues can be written as  tt t R p Q =⋅ where  t R is total revenues,  t p  is the weighted average 
price and  ti t i Qq ≡∑ , with  it q  being the quantity of good i  at time t . The total differential of the 
log of total revenues is: 
ln( ) ( / ) ( / ) tt t t dR p p Q Q =+ & &           ( 3 )  
where the dot stands for the time derivative. The average rate of change of prices is equal to the 
difference between the rate of change in total revenues and the rate of change in quantities. With a 
negatively sloped demand curve, the rate of change in quantities is negative, becoming positive in 6 
the LHS of (3). Therefore, setting  / tt QQ = 0 &
  will produce a conservative estimate of the euro 
changeover effect. Table 2 reports the estimate of the minimum value of the changeover effect 
calculated as 0.7 times the λ’s.  
[Table 2] 
The estimated menu costs range from 1 to 1.2 percentage points and the changeover effect amounts 
from 38 to 49% of the official inflation rate, suggesting that the euro changeover has had a 
detrimental impact on the competitiveness of the Italian economy at an aggregate level. These 
figures are in contrast with the evidence previously reported by the ECB (2003), where the reported 
estimates of the changeover effect ranged between 0.1 and 0.3 percentage points. 
3. Conclusions 
Using an empirically testable version of a staggered price model, we document that a euro 
changeover effect occurred in Italy. Our estimates show that indeed the fraction of firms keeping 
their prices unchanged decreased after the changeover. In turn, this implies a euro changeover effect 
of about 40% of the inflation rate. As for criticism of the published statistics made by consumers, 
although it is correct to claim that there was a changeover effect, this is taken into account by the 
official inflation figures. All in all, the euro changeover has exerted a negative impact on the 
competitiveness of the Italian economy. 
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Calvo G. (1983), Staggered Prices in a Utility Maximizing Framework, Journal of Monetary 
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Table 1. GMM estimation results 
 CPI  CED  GDPD  HCPI 
θ  
0.4698 
(0.0099) 
0.4978 
(0.0176) 
0.4399 
(0.0155) 
0.4896 
(0.0095) 
c θ  0.0773 
(0.0063) 
0.0735 
(0.0061) 
0.0914 
(0.0172) 
0.0738 
(0.0047) 
Wald test  [0.0000] [0.0000]  [0.0000]  [0.0000] 
Sargan J test  [0.6950] [0.5933] [0.7224] [0.5437] 
ξ   1.89 1.99 1.79 1.96 
c ξ  1.08 1.08 1.10 1.08 
() c
−− λ= ⋅ξ −ξ
11 4   1.59 1.69 1.40 1.66 
 
Note. The columns CPI, CED, GDPD, HCPI report the GMM estimation results for equation (2) in the main text where 
the inflation rate is calculated by using the Consumer Price Index, the consumption expenditure deflator, the GDP 
deflator and the Harmonized Consumer Price Index, respectively. The Wald test is for null hypothesis  : c H θ =θ 0 .  ξ  
and 
c ξ  are the quarters before price changes before and after the changeover, respectively.  λ  is the difference between 
the annual number of price revisions made by sellers between the changeover and the non changeover period. Standard 
errors in parenthesis.  p -values in square brackets. 
 10 
Table 2. Assessment of the euro changeover effect (1999q1 – 2003q4) 
 CPI  CED  GDPD  HCPI 
Average  inflation  2.39 2.66 2.55 2.36 
Menu  costs  1.11 1.19 0.98 1.16 
(Changeover  effect)  (46.47) (44.56) (38.52) (49.34) 
 
Note. See Table 1. Share of menu costs over the average actual inflation in parentheses. 

A Near Rational Model of the Business Cycle with Wage and Price Inertia,

Annual Report

Econometric Analysis, 6th Edition, Upper Saddle River,

Inflation Bias after the Euro: Evidence from the UK and Italy,

Inflation Dynamics: A Structural Econometric Analysis,

Long Term Contracts, Rational Expectations and the Optimal Money Supply Rule,

Managerial and Customer Cost of Price Adjustment: Direct Evidence from Industrial Markets,

Money Illusion and Coordination Failure,

Nominal Wage Contracts, Labor Adjustment Costs and the Business Cycle,

Staggered Prices in a Utility Maximizing Framework,

Staggered Wage Settings in a Macro Model, American Economic Review Papers and Proceedings,

Testing the Calvo Model of Sticky Prices, Economic Perspective,

The Magnitude of Menu Costs: Direct Evidence from Large US Supermarket Chains,

The Phillips Curve under State-dependent Pricing,

The Euro Changeover and Price Adjustments in Italy

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