Application of Concurrent Seasonal Adjustment to the Consumer Price 

Index1

Daniel Chow 

Adrian Thibodeau and Jeff Wilson 

Bureau of Labor Statistics 

2 Massachusetts Avenue, N.E., Room 3615, Washington, D.C. 20212, chow.daniel@bls.gov

Disclaimer: Any opinions expressed in this paper are those of the authors and do not constitute policy of the Bureau of 

Labor Statistics. 

1 We are grateful to Alex Stuckey of the Time Series Analysis Section of the ABS for his excellent explanations and

technical assistance, and to Craig McClaren and Mark Zhang for use of the ABS’ seasonal analysis program SEASABS, 

versions 2.4 and 2.5.1. Special thanks are extended to Jim Buszuwski, Claire Gallagher, and Stuart Scott at the BLS for their 

helpful suggestions, support, and feedback. 

Key words: seasonal adjustment, concurrent adjustment, seasonal revisions, forward factors, CPI, BLS, Australian 

Bureau of Statistics, SEASABS, X-12-ARIMA
GENERAL INTRODUCTION 

Background

The use of concurrent seasonal adjustment has received increasing attention from national and international statistical 

organizations for over 20 years. The long term trend toward greater demand for timely and accurate data, dramatically 

declining computation costs, and advances in statistical methodologies has led to a growing willingness to consider and adopt 

new methods of seasonal adjustment such as concurrent adjustment. Furthermore, the presence of established users of 

concurrent adjustment (Statistics Canada, US Census Bureau) and relatively recent users (US Bureau of Labor Statistics 

Current Employment Survey, Australian Bureau of Statistics Retail Trade Series) indicates that the benefits of switching to 

concurrent adjustment from the more established forward factors method are well understood by these organizations and 

many others worldwide. As the practice and understanding of concurrent adjustment increases, analysts will continue 

developing new methods to generate the most current and accurate seasonal data. In keeping with this growing trend, this 

paper seeks to evaluate the concurrent method for use on the US Bureau of Labor Statistics Consumer Price Index (CPI) and 

draws upon the approach developed by the Australian Bureau of Statistics (ABS) for its retail and business time series 

(Quarterly Economic Activity Survey (QEAS)). 

A major distinction between forward factor seasonal adjustment and concurrent adjustment is the use of the most recent 12-

months seasonal factors under the forward method as a proxy for current period seasonal factors. Concurrent adjustment uses 

the most recent data up to the current time period (month or quarter) to estimate the seasonal component. The former 

provides static estimates once per year and the latter creates accurate but less static estimates as new data are introduced each 

time period. Since 1996, the CPI has used the X-12-ARIMA forward factor approach to compute new seasonal factors to be 

applied to monthly unadjusted indexes in the upcoming new year. The estimates of the entire previous year’s factors are 

created and then revised with the next annual data, resulting in more stable backward revisions. This provides more 

conservative estimates based on longer term trends that are allowed to develop more fully before factor calculations are 

made. However, limiting the number of revisions using longer-span data may not be as precise in reflecting 

contemporaneous movements that can give earlier indications of trends. By more frequently updating estimates of trend, 

irregular, and seasonal components the concurrent adjustment method produces more dynamic and hence more accurate 

estimates, on average, of the final seasonally adjusted benchmark values of a time series. This frequent revising is an 

advantage because concurrent uses past data to generate up-to-date versions of seasonally adjusted values for the current time 

period. Every adjusted observation is thus more likely to converge and stabilize to its long run final values during subsequent 

estimates. 

This study uses the ABS’ approach outlined in “Review of Concurrent Adjustment for Retail and its Application to QEAS” 

(2001) using monthly data and a longer data span. It also draws upon some of the ABS’ prior work in its Information Paper, 

“Introduction of Concurrent Seasonal Adjustment into the Retail Trade Series” (1999). Readers should refer to these papers 

for further background details as some aspects of their applications are only summarily or briefly addressed here. The ABS 

work provides a unified framework, clear analytic observations, and a practical application of concurrent adjustment. Rather 

than focusing on concurrent adjustment theory, this analysis emphasizes the application of the concurrent method using 

simulations of monthly CPI data. It also relies on the SEASABS software which is an X-11(enhanced) based approach for 

adjusting time series data. 

Compared to its traditional forward factors method, the ABS finds “there are substantial gains in accuracy for concurrently 

adjusting quarterly series”(2001)2

 and demonstrates that the more frequent revisions compared to forward adjustment result 

in faster convergence to standard benchmark values for any given quarterly observation within a sub-span of concurrently run 

values. These results are expressed as average absolute revisions (AAR), or the mean absolute difference between the 

concurrent revisions and benchmark values versus the differences between forward revisions and benchmark at given lags of 

the time series. Graphs of the average revisions show clearly that percentage changes in revisions are fewer under concurrent 

adjustment compared to forward adjustment, with supporting evidence provided by measures of period-to-period movements, 

stability of trend and adjusted series rating (STAR), and level estimates of change. 

In its application of concurrent adjustment to higher level aggregate series, the ABS expected that: 

 

2 Mark Zhang and Andrew Sutcliffe “Use of ARIMA Models for Improving the Revisions of X-11 Seasonal Adjustment” 

Time Series Section, Australian Bureau of Statistics, November 2001, page 1 

1
the gains from using concurrent to be even greater for the disaggregated lower level series, where the data 

would be expected to contain a higher degree of volatility. The ABS has examined the performance of the 

concurrent methodology on many of the component series in the retail trade group and found that there were 

consistent gains for the group as a whole (2001).3

For the CPI, the stability and quality of lower level components is also important because these lower level components are 

adjusted before aggregation to US All items and other special aggregates. 

Another observation by the ABS is that concurrently adjusted quarterly data are more stable than monthly data, and that 

revision gains would be less substantial with monthly data (2001)4

. By using monthly CPI data for US level series, we can 

observe the extent of gains, if any, under concurrent adjustment and determine how monthly series exhibiting varying levels 

of volatility behave compared to the forward factor method. An additional feature of the study is based on the ABS’ 

observation “it is important to use a long data span when estimating average revisions so that the results will be 

representative” (2001)5

 for which the CPI analysis used 12-15 years of data, versus approximately 5 years in the QEAS study. 

Before CPI concurrent simulations are run, the test input series are analyzed using standard quality control diagnostics. 

Several stability measures (F(s), F(m), M7, and Q) are reported for each test component series after estimation in X-12-

ARIMA. These measures are used during the annual CPI seasonal analysis as guides for evaluating and maintaining the 

quality of the seasonal data. They are also applied to the concurrent simulations to identify similarities or differences in 

results produced during the most recent annual seasonal work. 

There are also several issues related to implementing concurrent adjustment that need to be examined, such as: how to use all 

currently available data to estimate seasonally adjusted trends in a monthly production process; determining how frequently 

revisions are to be made to previous estimates; setting policies regarding publication of concurrent data; establishing 

intervention analysis procedures; and developing seasonal adjustment diagnostics. Data quality, timeliness, transparency, 

and relevance are thus fundamentally important to the CPI program and to the data-using community. Identifying 

opportunities for improving methods and practices while establishing and remaining consistent with acceptable standards for 

concurrent adjustment standards is a worthy goal. This paper takes the first step in measuring the effects of estimating 

concurrently adjusted data using monthly CPI-U, US level time series indexes. 

The paper is divided into 6 major sections with an overview of key concepts in CPI seasonal adjustment followed by a 

discussion of concurrent seasonal adjustment background, an outline of the study’s design, the concurrent analysis results, a 

summary and discussion, and appendices. 

Seasonal Adjustment in the CPI

CPI seasonal factors are computed for the most recent past year and are applied to the upcoming 12 months of unadjusted 

data. While 8 years of data are used in seasonal computation, forward factors are revised for 4 years following an 

observation’s first year estimate. This results in a final value that is 5 years old and expected to be closer to the true seasonal 

trend. 

There are 73 seasonally adjusted components of the CPI-U, US All items index (see list in the Appendix A). The seasonal 

movement of the All-items index and other aggregations are derived by aggregating seasonally adjusted component indexes. 

Each January the seasonal status of every index series is reevaluated based upon statistical criteria. An index can change 

seasonal adjustment status from seasonally adjusted to non-seasonally adjusted, or vice versa. During mid-February each 

year when January data are released, revised seasonally adjusted indexes are published and new seasonal factors for the 

upcoming year for these items are available to data users. The CPI uses the U.S. Census Bureau’s X-12-ARIMA (autoregressive integrated moving average) seasonal adjustment software. X-12-ARIMA was developed as an improvement over 

the previously used X-11-ARIMA methodology. X-12-ARIMA uses the X-11 seasonal adjustment method in conjunction 

with regression-ARIMA modeling for intervention analysis and data extension. 

 

3

 Ibid, page 28 

4

 Ibid, page 30 

5

 Ibid, page 40 

2
Seasonal statuses of CPI series are evaluated each year, resulting in a decision to either “adjust” or “not adjust”. The X-12-

ARIMA procedure is applied to adjusted series. Non-adjusted series are combined with adjusted series and aggregated to 

estimate seasonally adjusted data for the higher level (i.e., indirectly adjusted) All-items, major groups, and special 

aggregates. Quality control diagnostics are used to monitor any residual seasonality which might exist in the irregular 

component of each series. 

 

Concurrent Adjustment Studies

Numerous studies have demonstrated the advantages of concurrent adjustment. Kenny and Durbin’s (1982) empirical study 

found “significant overall advantages” to concurrent adjustment over forward adjustment in improving local trend estimates. 

This work was supported by McKenzie (1984) who found greater general accuracy in estimates of the final level and monthto-month movements for Census Bureau data using X-11, as well as fewer extreme deviations from the final values. 

Shulman (1984) identified overall improvement in X-11 based concurrent trend estimations versus projected factors. Dagum 

and Morry (1984) noted the existing literature on the advantages of concurrent but added “the benefit of using concurrent 

seasonal factors is practically null, however, if seasonality is very regular or if the most recent values are strongly 

contaminated by outliers”. Dagum (1986) analyzed the consistency of forward and concurrent filters on revision frequencies 

and recommended combining the two revision methods. Pierce and McKenzie (1987) showed the theoretical expected gains 

from concurrent factors relative to forecast-augmented factors and to the amount of preliminary data error in the not-adjusted 

series. Bobbitt and Otto (1990) tested the improvement in X-11 forecasts via extensions to concurrently adjusted series and 

found these forecasts did no worse than ARIMA fitted models. 

Within BLS, a study of industry employment statistics by Kropf, Manning, Mueller, and Scott (2002) concluded that 

concurrently adjusted employment data are more accurate, leading to fewer revisions between the primary estimates and final 

benchmark series. They also observed that using concurrent adjustment would be advantageous during the Current 

Employment Survey (CES) program’s conversion to NAICS. Methee and McIntire (1987) found noticeable improvements in 

both the level and the month-to-month changes for employment/unemployment component series using concurrent 

adjustment. Buszuwski (1987) found that using concurrent adjustment combined with ARIMA forecasting reduced the mean 

absolute error of seasonal revisions relative to estimates without forecasting. 

In addition to consideration of concurrent adjustment, improvements and streamlined seasonal adjustment practices in the 

CPI have a long history. Development of BLS seasonal adjustment concepts and practices was first described in the 

published bulletin, “The BLS Seasonal Factor Method (1960)”, in which seasonal adjustment was initially applied to labor 

force series. CPI seasonal factor data first appeared in 1963 on a limited basis for 66 selected series from 1953-1961. This 

data was revised in the early 1960’s incorporating refinements in seasonal methodologies and in computer punch-card 

programming techniques. Furthermore, advances in mainframe computer technology in the 1970’s allowed faster and more 

systematic analysis of seasonal indexes. By 1980 the development of the Census Bureau’s X-11 program (and later versions) 

further raised processing speeds and allowed increasingly sophisticated calculations. In 1989-1990, the CPI began using the 

X-11-ARIMA and RAMP programs to automate and improve the quality of seasonal adjustments. In 1993 the CPI switched 

to the improved version of X-11-ARIMA program known as X-11-ARIMA/88. 

Official CPI adoption of the X-12-ARIMA standard in 1996 included development of a special graphical user interface 

(known as Vx12) plus several analytic support improvements introduced in subsequent years. In 2001 the CPI switched from

using the special seasonal aggregation weights that had been used since 1982 and began applying the biannually updated 

aggregation weights used to calculate the monthly unadjusted indexes. 

ANALYSIS DESIGN 

Component Series and Testing

In simulating the ABS method using monthly CPI-U data, several seasonally adjusted test series were chosen to provide 

enough variability in results by reflecting a mix of characteristics such as the degree of volatility, the number of 

interventions, and experience with the series during previous seasonal adjustments. The initial sample consisted of the 73 

seasonal components to All items, with a 9 series sample selected according to their weight and the characteristics above. 

The test series and corresponding item code identifiers are as follows: Owner’s equivalent rent of primary residence (SEHC), 

Motor fuel (SETB), Electricity (SEHF01), Video and audio (SERA), Women’s apparel (SEAC), Men and boys’ apparel 

(SAA1), Utility (piped) gas service (SEHF02), Cereals and bakery products (SAF111), and Dairy and related products 

(SEFJ). 

3
The simulation data span is 15 years plus 4 months from January 1990 to April 2004, a sufficient period for the movement 

patterns to fully develop in the unadjusted indexes. However, one series, Video and audio, was first calculated for the CPI in

January 1993 so only 12 years and 4 months of data were available. Concurrent simulation results for this shorter series were

compared to the longer-dated series and thus permitted some insight into the quality of concurrent adjustment resulting from 

a shorter span. Using 12-15 years of monthly data is a reasonable assumption based on the fact that concurrent results will be

more representative with longer lengths. Both the forward and concurrent estimation use these 12-15 year spans to generate 

benchmark estimates. Concurrent estimates are generated using a 4 year plus 4 month subspan (January 1997 to April 2001) 

within the benchmarks. 

Based on an enhanced version of X-11, SEASABS provides users with default options to automatically detect and fix 

specific data problems such as trend-breaks, trading day, and moving holiday and thus adjust for significant events specific to

Australia (eg, Australia Day, Chinese New Year, Easter, Father’s Day). None of these effects are required for CPI seasonal 

adjustment so the automatic options to apply the ABS corrective measures were not activated. 

Other options available in SEASABS allow adjustments similar to those under X-12-ARIMA based seasonal adjustment. 

The SEASABS test options include application of 3x5 final seasonal filters; using 13 term Henderson filters to final factors; 

seasonal decomposition via multiplicative adjustment; and applying the default sigma limits (1.50 lower, 2.50 upper) for 

weighting extreme values. The 9 test component series’ concurrent simulations consisted of simple SEASABS runs of noninterventioned 12-15 year data spans. These series are termed “Series1”. 

Stability Diagnostics

One indication of the quality of a concurrently adjusted series is the stability of final seasonally adjusted values and trend 

estimates. Seasonal stability in this paper is measured using diagnostic statistics calculated and reported along with outputs

from seasonal adjustment programs such as X11 and X-12-ARIMA. They are used to determine if a time series’ seasonality 

is present and identifiable in an adjusted series. 

The statistics routinely used in the CPI (F(s), F(m), M7, and Q) are explained in more detail here. 

F(s) is a test for the presence of seasonality assuming stability. Values greater than 7.0 indicate seasonality is present. 

Reported in Table D8 of X-12-ARIMA’s output file, this diagnostic is the most important one used during seasonal 

adjustment. 

F(m) is used to determine if the seasonal component is identifiable in the presence of moving seasonality. If moving 

seasonality is weak or not present, an acceptable F(m) value is 3 or less. It is reported in Table D8 of the X-12-ARIMA 

diagnostic output file. 

M7 is the ratio of the amount of stable seasonality to the amount of moving seasonality. There are eleven M-values produced 

by X-11 based programs each measuring a different type of problem in the data series. M7 carries the highest weight among 

them and is sensitive to the overall quality of seasonal adjustment. Values can range from 0 to 3 but values less than 1 

indicate the presence of identifiable seasonality. 

Q is a weighted average of the 11 M-statistics generated in X-11 based programs. It is a global measure of the quality of 

seasonal adjustment. Values less than 1 indicate a good overall adjustment. 

The AIC (Akaike’s Information Criterion) is normally used to compare the goodness of fit among different ARIMA models 

using the same time series. In this paper, if X-12-ARIMA is unable to automatically fit a model, no AIC value will be 

reported. If a series without an AIC value improves with an adjustment procedure, a value will be reported. SEASABS 

version 2.4 does not contain a fully integrated X-12-ARIMA program so AIC values are not reported for concurrently 

adjusted series. 

Stability is an important consideration in comparing concurrent and forward factor methods. Ideally, earlier observations 

should stabilize (i.e., not be subject to significant revisions) at some time point, while subsequent observations should 

converge quickly and smoothly to the final benchmark. We also expect series with lower STAR (Stability of Trend Analysis 

4
and Revision)6

 values, which measures the volatility of a time series, to have better stability statistics than series with higher 

STAR values. These are effects that can be tested through application of the concurrent method to monthly data. 

Stability of Test Series

Prior to simulating concurrent adjustment, an evaluation of the suitability of the Series1 test series is needed to determine if 

they are stable enough to provide meaningful concurrent output. Each one was run using X-12-ARIMA with the default 

specification files normally applied during CPI annual production. The following results compared to the desired thresholds 

were obtained: 

Table 1. F(s)>7 F(m)<3 M7<1 Q<1 AIC 

Series1.REQ.SEHC 15.483 3.183 0.731 0.39 31.6913 

Series1.MotorFuel.SETB 6.960 4.112 1.179 0.86 NA 

Series1.Electricity.SEHF01 414.124 6.466 0.179 0.20 475.3353 

Series1.VideoAudio.SERA 15.672 1.634 0.616 0.39 135.8315 

Series1.WomensApparel.SEAC 225.662 4.843 0.128 0.36 NA 

Series1.MenBoysApparel.SAA1 112.010 3.877 0.288 0.38 417.6443 

Series1.UtilityNatGas.SEHF02 10.538 2.539 0.833 0.63 NA 

Series1.Cereals.SAF111 30.055 1.469 0.436 0.26 278.6636 

Series1.Dairy.SEFJ 13.383 12.345 1.283 0.65 486.3780 

“NA”=Not Applicable; X-12-ARIMA did not choose an acceptable model for this series. 

In the CPI, all adjusted series are required to pass certain statistical diagnostics in order to be considered an acceptable 

adjustment. Table 1 column headings report the generally acceptable thresholds expected for seasonal stability. In some 

cases X-12-ARIMA did not identify acceptable ARIMA models and thus seasonally adjusted but did not provide AIC values 

for SETB, SEAC, and SEHF02. Interestingly, the results are generally acceptable when using the larger time span. F(m) 

results appears worrisome but F(m) is a minor diagnostic. F(s) is generally the most significant diagnostic showing total 

stable seasonality. M7 and Q also tend to be within acceptable ranges with some M7 results slightly above 1. Overall, using 

the longer-term, non-IASAed, unadjusted data will be reliable for SEASABS concurrent simulations. It is also expected that 

the problems encountered above will be mitigated to some degree by the more frequent revisions required as each month’s 

data are adjusted concurrently. 

Several of these series are subject to “intervention analysis for seasonal adjustment” to remove the effects of outliers and 

level shifts. For this paper, level shifts and outliers remain in the unadjusted test Series1 observations and are allowed to 

influence the concurrent simulation results. 

Testing and Extreme Outliers

The issue of how to treat outliers and level shifts in a concurrent system deserves attention. The presence of these volatile 

data movements introduces a degree of error and bias into the resulting seasonal factors and adjusted time series. 

Under the existing forward factors approach for the CPI, problematic outliers and level shifts are identified in the unadjusted

data spanning the previous eight years. The extreme observations are carefully selected and analyzed. When significant 

outliers and level shifts are identified by the analyst, the Census Bureau’s X-12-ARIMA software is used to estimate 

replacement values for the month(s) where the effects occur. The level shifts and outliers are removed from the original 

series prior to estimation of the seasonal factors. The seasonal factors are applied to the original series, and as a result the 

interventions remain in the seasonally adjusted data. 

Monthly identification and removal of level shifts and outliers is problematic because their true magnitude, duration, and 

direction are revealed with only the addition of subsequent data. Unlike the benefit of hindsight permitted under a forward 

factor method, no confirming evidence in the current month will likely exist to indicate how an outlier or level shift 

movement will evolve in future months. This leads to ad hoc assumptions about the identification and removal of 

outlier/level shift effects, particularly among volatile CPI series such as motor fuels. This in turn creates uncertainty about 

 

6

 See the next section, “The ABS and Concurrent Adjustment”. 

5
the appropriateness of choosing current and past outliers/level shifts, raising the potential for biased factor estimates at each 

concurrent revision. 

 

One solution is to allow extraneous effects to remain rather than guess the direction, timing, and mix of level shifts and 

outliers. For some series this may mean waiting until December data are available and conducting an annual analysis in 

January. Given the complexity of CPI seasonal adjustment, any problems accurately identifying level shifts and outliers 

under a concurrent process would present difficulties in producing reliable estimates. Thus, demonstrating the revision gains 

from seasonally adjusting the CPI concurrently is the primary focus of this study, with the treatment of outliers and level 

shifts reserved for treatment in the section, “Adjustment for Extreme Outliers and Level shifts”. Measuring the level of 

revision improvement from concurrent adjustment after the removal of level shifts is described in a later section. 

This study assumes frequent updating will reduce the effects of outliers and level shifts on seasonal time series. Concurrent 

estimation is tested without resorting to problematic short-term identification decisions that can introduce bias and errors. 

Permitting level shifts and outliers to remain in the series during concurrent simulations allows more consistent assessments 

of how well the SEASABS method operates using different CPI series. 

Summary

In determining the scope and level of simulations, several factors had to be considered. The quality of concurrent adjustment 

is a function of the frequency of revisions a given time series is subjected to, which is dependent on the number of 

observations. As more data are added, the X-11/X-12-ARIMA asymmetric filters become more accurate in extending the 

endpoints of the time series with forecasts and backcasts to estimate the final adjusted series. Shorter series tend to result in 

greater forecast errors, and thus greater error in estimates of trend and irregular components. As the number of observations 

grows, improved estimates of the seasonal patterns become apparent. 

This is a volatility-based study of initial test series representing the wider-ranging patterns found in the monthly CPI. Noisy 

data purportedly can be adjusted more accurately under concurrent compared to forward adjustment due to the higher number 

of revisions. Less noisy data should show little gains from revision. Thus, choosing series with various volatilities provides 

insights at the range of gains under a concurrent CPI system. Emphasis is placed on three areas: (1) comparative analysis of 

stability statistics, (2) simulating the average revisions between the concurrent and forward factor methods, and (3) 

discussion about outliers and level shifts. 

THE ABS AND CONCURRENT ADJUSTMENT 

Readers may refer to the ABS papers for a more complete discussion on the application of concurrent adjustment. A 

condensed summary of concepts necessary for further reading are provided here. Monthly data are assumed in this section 

although this concept can be applied to other frequencies. 

Average Absolute Revision

Several years of additional data from the initial introduction for a given month is needed for that observation to stabilize and 

thus be subject to very few revisions. This historically adjusted series is referred to as the “benchmark” estimate, or Xt,T , 

where X represents either the level or period movement of a series, t is a monthly point in time, and T is 3 years from the end

of series. However, an updated simplification by the ABS for this algorithm requires only that the entire span including T be 

run to produce the benchmark series. 

The concurrent method produces seasonally adjusted values using the data from the start of the span up to the latest

observation. Each time this is done for a new observation, prior monthly estimates are revised with new seasonally adjusted 

values. Each monthly concurrent update produces revised values that are more likely to converge to the benchmark. 

To compare forward and concurrent simulations, revisions against the benchmark for t are run for a sub-span of data between 

7 years after the first observation and 3 years (T) before the end of the series. Seven years are needed to produce reliable 

estimates and 3 years at the end of the series are isolated to keep the final estimates from being changed by revisions. The 

revision against the benchmark is interpreted as the total level of revision required for an estimate to reach the benchmark; 

i.e. the estimate for a particular time t after more than 3 years (T) of additional data becomes available. Thus, T represents

the total remaining revision from the concurrent estimate from k periods ago (or lag k) to the final estimate. 

6
The revision against the benchmark at a given month t is initiated by the addition of an unadjusted observation at each point 

in a defined lag k range. For example, lag k=0 indicates estimating the seasonally adjusted value up to and including the 

month of interest t; k=1 indicates the seasonal adjustment using the month ahead data from t; k=2 is the estimated value two 

months ahead of t, etc. At the end of lag k, the revision restarts with (t+1, k), (t+2, k),..., (T-t, k). The longer the value of k 

the more frequently the earlier or “lagged” data will be revised. 

For all values t there is a set of revised seasonally adjusted values. As more observations across t and k are used in the subspan of the benchmark, the frequency of revisions to observations increases, and hence the faster the convergence to 

benchmark. Estimates of the revision experience of a sub-span generated during a concurrent simulation will usually

measure the level of revision that can be expected for the entire time series. The earliest observations will be revised most 

often and not converge as quickly as later observations, but are more accurate compared to forward factors in those time 

periods. However, as newer data are received and because the information contained in them is more recent, the improved 

seasonal estimates converge more closely to the benchmark compared to forward factors. 

The percent revision for each observation X in period (t,T) and for lag k is defined below in Equation 1: 

( )

⎥

⎥

⎦

⎤

⎢

⎢

⎣

⎡ −

=

X

X X

R

Tt

Tt X kt

kt

,

,

,

,

ˆ

100

Where Xˆ

t,k

 is the estimate of the quantity of interest as either the level or period-to-period movement of the seasonally 

adjusted series at period t using all data up to t+k. Xt,T is the benchmark estimate for t using all data up to T. 

The Average Absolute Revision (AAR) is defined as measurement of the revisions at a specific lag k, in Equation 2: 

∑ =

=

e

st

X

kt

k

X

k R

n

R ,

1

where t=s is the start and t=e is the end of the simulated revision span, respectively, and nk is the number of observations in 

the simulated revision span at lag k. This formula measures the average absolute revision between the benchmark at time t 

versus the simulated estimates produced at time t for each k period observation in the future, for all values of t. Equation 2 is 

applied to both concurrent and forward factor methods to compare the rates of convergence to their respective benchmarks. 

The AAR can be depicted graphically to map the percentage movements and levels of convergence at each lag k. In the ABS 

example below, the dot points represent the AAR, to the level estimates, at specified lags for k=24. The blue and red lines 

are the AAR, to the level estimates, experienced under forward factor and concurrent analysis, respectively. At each lag, the 

concurrent point/line segment is lower than the forward factor’s, indicating faster convergence to benchmark and greater 

gains from seasonal revisions by switching to concurrent adjustment. 
Stability of Trend Analysis and Revision

In addition to the AAR, the ABS applies the Stability of Trend and Adjusted series Rating (STAR) developed by Sutcliffe 

(2000). It is a summary measure of the amount of average absolute percentage movement in the irregular component of an 

adjusted series, defined in Equation 3: 

∑= −

−

−

−

=

N

t

t

t t

I

I I

N

STAR

2 1

1

( )

100

1

1

where It

 is the irregular component at time t and N is the number of observations in the time series. The larger the STAR 

value, the greater the volatility and thus the greater expected gains from seasonal adjustment and trend revision. Series with

lower volatility will generally have lower STAR ratings, indicating that average revisions are smaller and convergence faster 

than relatively more volatile series. Although revision gains are expected to be greater for higher STAR values, it is more 

desirable that revisions are minimized at each update and that adjusted estimates converge quickly to final benchmark values. 

The ABS study notes that the STAR measure is a rough preliminary guide for the actual gains from concurrent, as each series 

varies in the amount of volatility and in the quality of seasonal adjustment when new data are added. Values greater than 10 

for monthly time series are considered volatile; however, volatility definitions will vary depending on the type of data series, 

how it is generated, its historical patterns, and the data users’ experience. Both X-11 and X-12-ARIMA estimate this value 

as an automated output statistic found in Table F3A, the first row of column D13-I. The degree of stability or instability for

time series cited in this paper is determined by the range of variation in the STAR values produced by the seasonal 

adjustment methods. 

ANALYSIS RESULTS 

Concurrent Test Series

Application of concurrent methodology using SEASABS on SERIES1 with a 15+ year span of data is compared to results of 

the original X-12-ARIMA forward factors results calculated using 5-8 years of data as is done in CPI production. These X12-ARIMA forward seasonal adjustments are termed “Standard Benchmark” (StdBench). The goal of this work is to 

determine if concurrent adjustment compares favorably to the standard CPI adjustment methodology. By contrast, for the 

ABS papers and AAR graphs, “standard benchmark” refers to the X-11 adjustment of the entire data span (typically 15 years 

in this study) to measure the rates of convergence between the forward and the concurrent methods. The standard benchmark 

for concurrent simulation purposes is the seasonal adjustment using the entire dataspan of each time series. 

Quality control statistics and Annual Average Revisions (AAR) graphs for seasonal factors are reported for each simulation. 

Note that the concurrent runs, which include the forward factor simulation, can only be applied to directly adjusted series. 

1
This is because forward factors cannot be simulated for indirectly adjusted series. Thus, the following examples are direct 

adjustments only. 

Dairy and related products Example 

Series1 concurrent adjustment uses a 15 year plus 4 month span, versus 8 years for the X-12-ARIMA forward adjustment, 

Dairy(StdBench). Concurrent adjustment should result in increased overall gains, on average, to seasonal factors compared 

to the forward factor method. Contrary to expectation of divergent results between StdBench and Series1, the results here 

demonstrate consistent gains in both the simulated concurrent method and the status quo forward factor method: 

Table 3a. STAR F(s)>7 F(m)<3 M7<1 Q<1 AIC 

Dairy, Series1 0.37 12.593 10.612 1.240 0.92 **** 

Dairy (StdBench) 0.32 14.578 10.206 1.136 0.58 305.1596 

  Not available. 

Both results are generally unstable with high F(m) and M7 values, although DairySeries1 is marginally more unstable than 

Dairy(StdBench). The STAR values, while very low (clearly less than 1), is highest for Series1, indicate slightly worse

volatility. F(m) is quite high in both cases. This indicates that high levels of seasonality remain undetected by the two 

methods, compared to the fairly stable F(s) values that indicate some acceptable stability in the series. The time series is a

candidate for further investigation and work, but the results are not inconsistent. 

Graphically (Figure 1.1), Series1 concurrent adjustment converges faster on average than its forward factor method. The 

blue forward factor AAR line of Figure 1.1 is the expected convergence path for Dairy(StdBench) using the longer span in X11 and other assumptions of this study. The positive gains from concurrent are the areas of vertical difference in percentage 

revision (PCT REV) between the blue (forward) and red (concurrent) at the lines and points at every lag. DairySeries1 shows 

that the concurrent AAR converges overall faster than their forward AARs, thus producing positive gains from revisions 

under concurrent adjustment at each lag. 

The decline of revision efficiency at lags 12 and 24 occurs when a second observation of the same month is available and a 

better estimate using the symmetric seasonal moving average is used (see ABS (2001) technical notes). 

Figure 1.1 Dairy Series1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

DAIRY.SEFJ

Pct rev

 0.1250

 0.1500

 0.1750

 0.2000

 0.2250

 0.2500

 0.2750

 0.3000

 0.3250

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

1
Electricity Example 

For Electricity (SEHF01), a highly seasonal series with interventions, the results are clearer: 

Table 3b. STAR F(s)>7 F(m)<3 M7<1 Q<1 AIC 

Electricity, Series1 0.41 422.364 6.061 0.173 0.22 ** ** 

Electricity (StdBench) 0.28 329.513 5.560 0.190 0.22 260.9612 

  Not available. 

As in Dairy, ElectricitySeries1 has the higher relative STAR value compared to Electricity(StdBench), indicating greater 

volatility in the longer span. Series1 has the greater acceptable F(s) and a larger F(m) compared to StdBench. In actual CPI 

production, significant level shifts are identified and removed for this series so we do not know beforehand to what extent 

revision improvements for concurrent estimation would be achieved if IASA was applied. Graphically, concurrent Series1 

AAR is superior to forward factors. 

Figure 1.3 Electricity SERIES1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

ELECTRICITY.SEHF01

Pct rev

 0.1000

 0.1250

 0.1500

 0.1750

 0.2000

 0.2250

 0.2500

 0.2750

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

Motor fuel Example 

The third IASA series, Motor fuel (SETB), is a more volatile series. 

Table 3c. STAR F(s)>7 F(m)<3 M7<1 Q<1 AIC 

Motor Fuel, Series1 2.03 7.541 3.836 1.108 0.98 ** ** 

Motor Fuel (StdBench) 1.14 6.355 4.043 1.227 0.91 ** 

  Not available. ** No Model Chosen in X-12-ARIMA. 

Series1 is a marginal improvement over Motor Fuel(StdBench) although both experience slightly less acceptable F(m) and 

M7 values and borderline Q. However, X-12-ARIMA was unable to identify an acceptable model for Motor Fuel(StdBench), 

indicating a high degree of volatility in the the unadjusted series. Series1 should converge faster on average than Motor 

Fuel(StdBench) based on Series1’s larger STAR value, which roughly predicts larger gains in the level of revisions the

greater the volatility of a time series. 

2
Figure 1.5 Motor Fuel SERIES1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

MOTORFUEL.SETB

Pct rev

 0.6000

 0.8000

 1.0000

 1.2000

 1.4000

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

In Figure 1.5, concurrent revisions again show overall better convergence than forward revisions. However, deterioration in 

the AAR occurs at lags 8-11 following a period of more smoothly declining convergence. Long-lived adjustment effects at 

those lags deteriorate the quality of AAR revisions before returning to its downward trend slope. The revision point at lag 11

touches but does not exceed the forward factor revision, indicating that extreme adjustment effects are on average modest. 

Despite the low stability, this series experiences a large average distance of gains in PCT REV between lags 0 and 24 of 

approximately 0.5 percentage point (1.25 minus 0.75). This is preliminary evidence of gains from concurrent adjustment 

helping to achieve faster and larger revisions than forward adjustment for volatile time series. 

At longer lags, Series1 settles near a PCT REV range of about 0.7 or 0.8. The stable slopes of revision improvement at these 

longer lags (20-24) suggest the percentage concurrent revisions rate of change are most rapid in the earlier lags 0-7 and at 

lags 13-20. However, the greatest average gain versus forward factors as shown by the vertical distances between the 

pathways is concentrated in lags 14-22 at varying degrees due to the less stable forward revisions. The greatest overall 

revisions are thus still achieved in concurrent adjustment than in forward adjustment. 

Utility (piped) gas service Example 

Unlike the previous examples, Utility (Piped) Natural Gas’ Series1 results are generally more acceptable than StdBench

while the difference in STAR scores is not excessive. Its F(m) value is just below the acceptance boundary and F(s) is more 

than twice that of Utility Nat Gas(StdBench) indicating increased stability from the adjustment, although StdBench’s F(m) is 

more acceptable. M7 and Q are also improved, particularly for the M7 ratio which represents whether identifiable 

seasonality is present. 

Table 3d. STAR F(s)>7 F(m)<3 M7<1 Q<1 AIC 

Utility Nat Gas, Series 1 1.17 11.361 2.751 0.819 0.80 ****

Utility Nat Gas (StdBench) 1.05 5.276 1.269 1.012 0.77 500.7268 

  Not Applicable. 

3
Figure 1.7 Utility Natural Gas SERIES1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

UTILITYNATGAS.SEHF02

Pct rev

 0.3000

 0.4000

 0.5000

 0.6000

 0.7000

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

Despite generally better diagnostics, Figure 1.7 shows Series1 produces problematic mixed average absolute revisions. At 

some lags forward adjustment converges faster than concurrent, and vice versa. Visual inspection reveals that forward 

adjustment is on average better than concurrent, with cyclical peaks and troughs at certain lags. Concurrent AAR improves 

initially at lags 0-1 then shows no appreciable declines until lags 12-13. By comparison, forward AAR revisions generally 

decline in trend but with decreasing amplitude. The mixed effects are due to volatility during the past five years in the 

unadjusted series making identification of the series’ components difficult. Utility natural gas is normally subject to IASA 

procedures to remove known significant extraneous movements. Hypothetically if its volatility could be accounted for by 

removal of level shifts and outliers prior to simulation then we should see further improvement in its diagnostics. Without 

empirical testing however, it is speculative to conclude whether or not such improvement leads to concurrent revision gains 

over the forward factor case. The AAR trends and lack of clear support for either approach suggests further investigation is 

needed. 

Remaining Test Series Examples 

The five remaining CPI test series— Cereals and bakery products (SAF111); Men and boys’ apparel (SAA1); Women’s 

apparel (SEAC); Owners’ Equivalent rent of primary residence (SEHC) or “REQ”; and Video and audio (SERA)—are 

components to seasonally adjusted All items. They illustrate several types of convergence profiles that may be more or less 

typical depending on the data span, whether seasonality is identifiable, and the degree of stability or instability. 

Table 3e. STAR F(s)>7 F(m)<3 M7<1 Q<1 AIC 

Cereals, Series1 0.19 28.606 1.316 0.440 0.32 ** ** 

Cereals (StdBench) 0.19 20.109 2.040 0.571 0.47 172.0122 

MenBoysApparel, Series1 0.39 119.039 3.799 0.280 0.36   

MenBoysApparel (StdBench) 0.44 71.403 4.666 0.383 0.39 242.7464 

Women’s Apparel, Series1 0.59 248.83 5.312 0.210 0.35 ** ** 

Women’s Apparel (StdBench) 0.56 145.585 5.468 0.248 0.31 ** 

REQ, Series1 0.06 14.070 5.309 0.900 0.48   

REQ (StdBench) 0.03 21.756 3.143 0.614 0.51 -56.0527 

Video and Audio, Series1 0.20 15.119 1.140 0.590 0.60 ** ** 

Video and Audio (StdBench) 0.17 15.215 0.464 0.525 0.38 70.5217 

  Calculated but not reported in SEASABS. ** No Model Chosen in X-12-ARIMA. 

4
Except for Owners’ Equivalent rent of primary residence and Video and audio, Series1 compares favorably overall against 

StdBench. F(s) values for Cereals, Men and boys’ apparel, and Women’s apparel are greater than StdBench. Series1 F(m), 

M7, and Q (except for Women’s apparel) are generally lower than their counterparts. STAR values are at the general levels 

of StdBench. Note that REQ, a very stable trend series, has a very low STAR measure. Series1 in this case does not improve 

much over REQ (StdBench). Cereals, another very stable series but only slightly less so than REQ, has identical STAR 

scores but Series1 is an improvement over Cereals (StdBench). 

Despite the other indicators, F(m) is above desired levels in Men and boys apparel, Women’s apparel, and REQ, although the 

more important F(s) is acceptable for all Series1. Like the IASA series, STAR values here are also generally low, indicating 

that a value near or above 1.0 is likely a more accurate measure of high volatility in CPI-U series. Low M7 and Q scores in 

Table 3e also indicate generally similar estimates of moving and stable seasonality. 

Series1’s diagnostics show that its adjustments are better than or similar to their StdBench series. Only Video and audio 

(StdBench) and REQ (StdBench) appear to show a better quality adjustment over Series1. Series1 F(s) values are generally 

greater. One explanation is that concurrently adjusted Series1 simulations’ longer (12-15 years) data spans are an advantage 

over the current forward factor method, which for StdBench are adjusted using 8 years of data, the minimum span needed. 

Another explanation is that the more frequent revision from updating each seasonal adjustment using new observations 

allows more accurate identification of the trend, irregular, and seasonal components relative to their long run values. The 

more frequent concurrent updates generate minor instability as indicated by the slightly high F(m) values, but creates more

accurate seasonal factor estimates during each revision. 

Graphical analysis reveals concurrent converges faster than forward factors to simulation benchmarks in the majority of 

cases. 

Figure 1.9 Cereals Series1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

CEREALS.SAF111

Pct rev

 0.0300

 0.0400

 0.0500

 0.0600

 0.0700

 0.0800

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

5
Figure 2.0 Men and Boys Apparel Series1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

MENBOYSAPPAREL.SAA1

Pct rev

 0.1000

 0.1250

 0.1500

 0.1750

 0.2000

 0.2250

 0.2500

 0.2750

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

Figure 2.1 Women’s Apparel Series1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

WOMENSAPPAREL.SEAC

Pct rev

 0.2000

 0.3000

 0.4000

 0.5000

 0.6000

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

6
Figure 2.2 Owners’ Equivalent Rent Series1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

REQ.SEHC

Pct rev

 0.0150

 0.0200

 0.0250

 0.0300

 0.0350

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24

Table Lags

Concurrent Forward Factor

Figure 2.3 Video and Audio Series1, Concurrent versus Forward Factor AAR 

Seasonal Factors : Average revision at specified lags.

VIDEOAUDIO.SERA

Pct rev

 0.0000

 0.0200

 0.0400

 0.0600

 0.0800

0 1 2 3 4 5 6 7 8 9 10 11 12 13 14 15

Table Lags

Concurrent Forward Factor

Summary

Concurrent AAR graphs can vary widely relative to forward factor AAR, but the majority of cases tend to experience large 

revision gains in the initial lags. This suggests that positive gains in initial lags provide early indications of the gains that 

likely can be expected in later lags. Early or mixed losses in concurrent AAR, as in the Video and audio example, will likely 

indicate no practical comparative advantage of concurrent over the forward factor revisions. However, exceptions such as the 

Utility natural gas example illustrate that some early initial gains may not always lead to a clear advantage to either revision 

scheme. Studying how well span length and the degree of stability (or instability) in all seasonal CPI component series affect

the performance of their concurrent and forward factor revisions would yield more precise information needed to describe 

their expected revisions history under a concurrent system. This would also help identify all specific cases where neither 

concurrent nor forward adjustment yields the greatest advantage and identify in what manner these cases are (or are not) 

problematic for seasonal factor estimates. This initial study using the SEASABS method of concurrent adjustment on CPI 

7
data shows that some unique and special cases do exist but are less common than cases where concurrent average absolute 

revisions are less than under the forward case. 

Adjustment for Extreme Outliers and Level Shifts

This section focuses generally on the mechanics of adjusting for level shifts and outliers rather than on specific data

simulations. This section serves as a guide for further investigation on this topic. 

The Intervention Analysis Seasonal Adjustment (IASA) technique is used in the seasonal adjustment of the CPI to provide 

more accurate data. This process offsets the effects that extreme price volatility would otherwise have on the estimates of 

seasonally adjusted data. Intervention analysis is the prior adjustment of an index series before the calculation of the seasonal 

factors. Prior adjustment may be called for if an "outlier” or “level shift" occurs. (A "level shift" occurs when a good or 

service undergoes a unique, large, and rapid change in price level.) An example would be a large decrease in the price of 

gasoline due to the breakdown of an oil cartel. Removal of the level shift gives a clearer seasonal pattern and lessens the 

irregular component. When a level shift exists, intervention analysis helps to calculate more accurate seasonal adjustment 

data. Previous and current level shifts are published each January in the U.S. Department of Labor’s January “CPI Detailed 

Report” http://www.stats.bls.gov/cpi/cpisaia04.pdf. 

This study contains simulated concurrent adjustments of several higher-weighted component indexes to the CPI-U All items 

level. A few of these test series received intervention analysis to produce forward factors during the most recent seasonal 

production cycle. 

The goal of the IASA process is to estimate movement due to unusual non-seasonal economic events (interventions). The 

resulting seasonal factors, which better represent the seasonality of the series, are then applied to the original series. As 

mentioned earlier, specifying level shifts and outliers presents challenges for concurrent estimation. Identifying and

removing level shifts and outliers on a month-to-month basis in the CPI would likely result in identification errors. In 

practice, identification of interventions improves as time passes. In some cases, unusual movements can be identified 

immediately and the cause of the movement is widely known; in most cases, more is known about the cause of the movement 

months later. 

In order to simplify analysis, CPI test series in this study have no IASA procedure applied to them prior to estimation in 

SEASABS. For volatile series, where IASA procedures are normally employed, concurrent adjustment resulted in better and 

faster converging revisions estimates to the benchmark than forward factor adjustment without IASA. We hypothesize that 

for these series, concurrent adjustment combined with IASA procedures will result in better AAR results. 

Using level shifts and outliers in concurrent simulations was investigated but developing a reliable methodology was beyond 

the time allotted for this study. One approach was to remove the known level shift and outlier effects from extended-span 

versions of actual CPI series receiving IASA treatment in 2004 based on events reported in the January 2005 “CPI Detailed 

Report”. Longer-span data estimated in X-12-ARIMA yield the component series of B1 prior adjustments for extreme 

movements based on user-supplied variables for level shifts, and outliers. After selecting these B1 data (termed SERIES2) 

corresponding to their Series1 non-IASA counterpart series, they were used in the X-11 based SEASABS concurrent versus 

forward factor revisions simulations to produce AAR graphs, STAR values, and stability statistics. Not surprisingly, the 

concurrent AAR results obtained tended to improve over their forward factor revisions. 

A further challenge is the fact that data in this study is used and produced from two separate systems: (1) the X-12-ARIMA 

method using a CPI-designed GUI (called VX12) to facilitate annual IASA work but which has no concurrent estimation 

capability; and (2) the X-11 based simulations in SEASABS designed for concurrent and forward revision analysis but which 

does not provide flexible and efficient IASA capabilities found in VX12 for X-12-ARIMA. Further discussion can be found 

in Appendix B. 

CONCLUSION 

In this paper we see that revision gains from concurrent seasonal adjustment are possible for a variety of CPI series. 

Although these are simulations, the lagged revisions changes in AAR and stability statistics for non-IASA series show 

patterns consistent with those described by the ABS for quarterly series. We also find that in the case of monthly data the 

CPI series tend to show a wide range of revision gains relative to forward factors. The two important determinants in 

8
whether the gains are positive, negative, or mixed appear to be the length of the data span used in the concurrent estimation 

and the volatility present in the series. Given adequate data spans of over 15 years in some cases, which allows enough time 

for more frequent updates and thus allows lagged revisions to become more stable, uncertainty then lies in the degree to 

which seasonality is masked by volatility or noise. We have seen that most concurrent test series are consistent with the 

usual standard benchmark measures for the annual seasonal adjustment. Most test series also experience larger and faster 

revision gains than their forward factor counterparts, although some instability of concurrent revisions may be expected. The 

few test series with mixed revision gains suggest further analysis is necessary to determine whether such series can be 

addressed naturally by the addition of future observations (increased seasonal stability) or by the removal of outliers and level 

shifts. While the study does not explicitly discuss establishing a revisions policy, this also needs to be carefully considered in 

light of these findings. 

The presence of outliers and level shifts complicates the analysis. Comparison of test series with intervention effects 

included against counterpart series with intervention effects excluded requires that the original indexes be adjusted for the 

removal of outliers and level shifts in the B1 values whose seasonal adjustments are estimated in SEASABS. A general 

process is illustrated in Appendix B that may achieve this. 

Advances in statistical methodology continue to produce competing seasonal adjustment processes, particularly in widely 

used platforms such as TRAMO/SEATS, DEMETRA, and X-12-ARIMA, as well as in seasonal adjustment practices such as 

revision policies, model selection, and seasonal diagnostics. Concurrent adjustment is one well known alternative that 

presents the possibility for more timely data. It is more processing intensive than forward factors and may require more 

frequent monitoring for quality control purposes. 

Further study and research is encouraged to address other issues in CPI concurrent adjustment. 

9
APPENDIX A 

February 18, 2005 

Consumer Price Index: Components for Seasonal Aggregation to All items 

NSA = Not Seasonally Adjusted 

 

Item Title Status 

1 SAF111 Cereals and bakery products 

2 SEFC Beef and veal NSA 

3 SEFD Pork 

4 SEFE Other meats NSA 

5 SEFF Poultry NSA 

6 SEFG Fish and seafood NSA 

7 SEFH Eggs 

8 SEFJ Dairy and related products 

9 SEFK Fresh fruits 

10 SEFL Fresh vegetables 

11 SEFM Processed fruits and vegetables 

12 SAF114 Nonalcoholic beverages and beverage materials 

13 SEFR Sugar and sweets 

14 SEFS Fats and oils 

15 SEFT Other foods NSA 

16 SEFV Food away from home NSA 

17 SAF116 Alcoholic beverages NSA 

18 SEHA Rent of primary residence 

19 SEHB01 Housing at school excluding board 

20 SEHB02 Other lodging away from home including hotels and motels 

21 SEHC Owners' equivalent rent of primary residence 

22 SEHD Tenants' and household insurance NSA 

23 SEHE Fuel oil and other fuels 

24 SEHF01 Electricity 

25 SEHF02 Utility (piped) gas service 

26 SEHG01 Water and sewerage maintenance 

27 SEHG02 Garbage and trash collection NSA 

28 SAH3 Household furnishings and operations 

29 SAA1 Men's and boy's apparel 

30 SEAC Women's apparel 

31 SEAD Girls' apparel 

32 SEAE Footwear 

33 SEAF Infants' and toddlers' apparel 

34 SEAG Jewelry and watches 

35 SETA01 New vehicles 

36 SETA02 Used cars and trucks NSA 

37 SETA03 Leased cars and trucks1 NSA 

38 SETA04 Car and truck rental 

39 SETA09 Unsampled new and used motor vehicles1 NSA 

40 SETB Motor fuel 

41 SETC Motor vehicle parts and equipment NSA 

42 SETD Motor vehicle maintenance and repair 

43 SETE Motor vehicle insurance 

10
44 SETF Motor vehicle fees NSA 

45 SETG Public transportation 

46 SAM1 Medical care commodities 

47 SAM2 Medical care services 

48 SERA Video and audio 

49 SERB01 Pets and pet products NSA 

50 SERB02 Pet services including veterinary 

51 SERC Sporting goods NSA 

52 SERD01 Photographic equipment and supplies NSA 

53 SERD02 Photographers and film processing NSA 

54 SERD09 Unsampled photography1 NSA

55 SERE01 Toys NSA

56 SERE02 Sewing machines, fabric, and supplies NSA 

57 SERE03 Music instruments and accessories NSA 

58 SERE09 Unsampled recreational commodities1 NSA 

59 SERF01 Club membership dues and fees for participant sports 

60 SERF02 Admissions 

61 SERF03 Fees for lessons or instructions 

62 SERF09 Unsampled recreation services1 NSA 

63 SERG Recreational reading materials NSA 

64 SEEA Educational books and supplies 

65 SEEB Tuition, other school fees, and childcare 

66 SEEC01 Postage NSA

67 SEEC02 Delivery services 

68 SEED01 Land-line telephone services, local charges NSA 

69 SEED02 Land-line telephone services, long distance charges NSA 

70 SEED03 Wireless telephone services NSA 

71 SEEE Information technology, hardware and services NSA 

72 SEGA Tobacco and smoking products NSA 

73 SAG1 Personal care 

1This series is not published, but contributes its weight in the calculation of the All items index.

11
APPENDIX B 

Integration of IASA prior adjustments are problematic for comparing Series1 and Series2 because using prior adjusted (B1) 

data as input into SEASABS for concurrent adjustment results in correct seasonal factors which are then applied to the prior 

adjusted series, instead of to the original series as is desired. As a result, the seasonally adjusted data will not contain the 

level shift and outlier movements that are found in the original time series. To accurately measure AAR results between 

IASA (X-12-ARIMA based) versions and non-IASA (SEASABS, X-11 based) versions of CPI test series, it is necessary to 

restore the intervention variables removed from the B1 series simulations and apply them to the resulting seasonally adjusted 

series before performing statistical tests on the seasonally adjusted data. Both systems generate data used in the steps 

outlined below. 

Step1 

For each CPI series with IASA treatment applied to it during the 2004 CPI seasonal adjustment, run the seasonal adjustment 

in X-12-ARIMA using the same outliers/level shifts with the extended data span (15 years plus 4 months) instead of the 8 

years used in CPI seasonal adjustment. Obtain the B1 priors and run it in SEASABS concurrent simulation (Series2). 

Step2 

Our goal is to obtain the following, assuming monthly time series: 

(1) SeasAbs(SA) * PF = OrigSA 

where, 

• SeasAbs(SA) is the set of monthly seasonally adjusted indexes produced using the SEASABS program on the B1 

priors from Step1 (aka, Series2) 

• PF is the Prior Factors (Original Index / B1) derived from X-12-ARIMA in Step1 

• OrigSA is the desired time series originally estimated in X-12-ARIMA and containing the level shifts and outliers in 

PF. 

Step3 

Alternatively, (1) and (2) are equivalent: 

(2) OrigIX / SF(Abs) = OrigSA 

where, 

• OrigIX is the set of monthly unadjusted indexes used for X-12-ARIMA seasonal adjustment in Step1 

• SF(Abs) is the Series2 factors produced in SEASABS from B1 priors as the input data 

• OrigSA is the desired time series originally estimated in X-12-ARIMA and containing the level shifts and outliers. 

Step4 

Transform (2) as, 

(3) (OrigIX / PF) * (1 / SF(Abs)) = OrigSA / PF 

Let (3)=(4), where 

(4) SeasAbs(SA) = OrigSA / PF 

(5) SeasAbs(SA) * PF = OrigSA 

12
Step5 

Run OrigSA in SEASABS (SERIES3) and do the concurrent simulation versus forward factors. Compare Series3 (OrigSA, 

IASA-adjusted) to Series1 (non-IASA test series). 

As shown above, (1) is equivalent to (5). Both methods shown in (1) and (2) result in the SEASABS seasonally adjusted 

indexes (using B1 as input from X-12-ARIMA with IASA treatment) to be multiplied by PF (the ratio of Original Indexes to 

B1 priors derived from the extended span simulation and IASA application). The B1-based SEASABS seasonal factors 

should reflect the IASA effects removed. It is used to adjust the original unadjusted data (OrigIX / SF(Abs) = OrigSA) for all

user-specified level shifts identified and published as IASA events during the annual seasonal adjustment procedure. OrigSA 

should be equivalent to the SEASABS seasonal indexes (B1 itself seasonally adjusted) multiplied by the prior factors (PF). 

13
REFERENCES 

Material provided by the Australian Bureau of Statistics:

Dennis Trewin “Information Paper: Introduction of Concurrent Seasonal Adjustment into the Retail Trade Series” Australian 

Bureau of Statistics, December 1999 

Mark Zhang and Andrew Sutcliffe “Use of ARIMA Models for Improving the Revisions of X-11 Seasonal Adjustment” 

Time Series Section, Australian Bureau of Statistics, November 2001 

Nick Von Sanden and Mark Zhang “Review of Concurrent Adjustment for Retail and Its Application to QEAS” Time Series 

Section, Australian Bureau of Statistics, June 2001 

Australian Bureau of Statistics, SEASABS v2.4 and v2.5.1, 2004 

Additional papers:

David A. Pierce and Sandra K. McKenzie “On Concurrent Seasonal Adjustment” Journal of the American Statistical 

Association, Volume 82, Issue 399, September 1987 

David F. Findley and Brian C. Monsell “Using Quarterly Seasonal Adjustments to Improve Monthly Estimates” U.S. Bureau 

of the Census, Statistical Research Division, October 1993 

Estela Bee Dagum “Monthly Versus Annual Revision of Concurrent Seasonally Adjusted Series” Time Series Research and 

Analysis Division, Methodology Branch, Statistics Canada, 1986 

Estela Bee Dagum and Marietta Morry “Basic Issues on the Seasonal Adjustment of the Canadian Consumer Price Index” 

Time Series Research and Analysis Division, Statistics Canada. Journal of Business and Economic Statistics, Volume 2, 

Number 3, July 1984 

George R. Methee and Robert J. McIntire “An Evaluation of Concurrent Seasonal Adjustment for the Major Labor Force 

Series” U.S. Bureau of Labor Statistics. Proceedings of the Business and Economics Section, American Economic 

Association, 1987 

“Harmonising Seasonal Adjustment Methods in European Union and OECD Countries” OECD Short-Term Economic 

Statistics Expert Group, Organisation for Economic Co-operation and Development, June 2002 

Holly B. Shulman “Trend Estimation with Concurrent Seasonal Adjustment” U.S. Bureau of the Census, Statistical Research 

Division, 1984 

James A. Buszuwski “Alternative ARIMA Forecasting Horizons When Seasonally Adjusting Producer Price Index Data with 

X-11-ARIMA in Concurrent Mode” U.S. Bureau of Labor Statistics. Proceedings of the Business and Economics Section, 

American Economic Association, 1987 

James A. Buszuwski and Stuart Scott “Some Issues in Seasonal Adjustment When Modeling Interventions” U.S. Bureau of 

Labor Statistics. Proceedings of the Section on Survey Research Methods, American Statistical Association, 1993

Jurgen Kropf, Christopher Manning, Kirk Mueller, and Stuart Scott “Concurrent Seasonal Adjustment for Industry 

Employment Statistics” U.S. Bureau of Labor Statistics, October 2002 

Larry Bobbitt and Mark Otto “Effects of Forecasts on the Revisions of Concurrent Seasonally Adjusted Data Using the X-11 

Seasonal Adjustment Procedure” U.S. Bureau of the Census, 1990 

14
Penny Barber and Chris Zangel “The Treatment of Seasonal Commodities in the Consumers Price Index” Consumers Price 

Index Revision Advisory Committee, Statistics New Zealand, 1997 

P. B. Kenny and J. Durbin “Local Trend Estimation and Seasonal Adjustment of Economic and Social Time Series” Journal 

of the Royal Statistical Society, Series A (General), Volume 145, Number 1, pages 1-41, 1982 

Ronny Nilson “Seasonal Adjustment Methods in the European Union and Other OECD Countries” Organisation for 

Economic Co-operation and Development, The Statistics Newsletter for the Extended OECD Statistical Network, Issue 

Number 5, November 2001 

Sandra K. McKenzie “Concurrent Seasonal Adjustment with Census X-11” U.S. Bureau of the Census, Statistical Research 

Division. Journal of Business and Economic Statistics, Volume 2, Number 3, July 1984 

“Seasonal Adjustment Methods: A Comparison” Eurostat, September 1998 

“Seasonal Factors. Consumer Price Index: Selected Series June 1953-May 1961” U.S. Bureau of Labor Statistics, Bulletin 

Number 1366, May 1963 

Stuart Scott “On the Impact of Outliers on Seasonal Adjustment” U.S. Bureau of Labor Statistics. Proceedings of the 

Business and Economics Section, American Statistical Association, 1987 

“The BLS Seasonal Factor Method (1964)” U.S. Bureau of Labor Statistics, 1964 

ONLINE REFERENCES 

Information about the Census Bureau’s X-12-ARIMA software and supporting documentation are available via the internet: 

http://www.census.gov/srd/www/x12a/x12down_pc.html

Further information about the Bureau of Labor Statistics CPI seasonal adjustment methods and data are available via the 

internet: 

CPI seasonal adjustment 

http://www.stats.bls.gov/cpi/cpisapage.htm

U.S. Department of Labor, January “CPI Detailed Report” 

http://www.stats.bls.gov/cpi/cpisaia04.pdf

Main CPI web page 

http://www.bls.gov/cpi/home.htm

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