An Empirical Decomposition of Risk and Liquidity in

Nominal and Ináation-Indexed Government Bonds

Carolin E. Páueger and Luis M. Viceira1

First draft: July 2010

This version: May 2011

1Páueger: Harvard Business School, Boston MA 02163. Email cpáueger@hbs.edu. Viceira: Harvard

Business School, Boston MA 02163 and NBER. Email lviceira@hbs.edu. We are grateful to John Campbell, Graig Fantuzzi, Josh Gottlieb, Robin Greenwood, Arvind Krishnamurthy, Jeremy Stein, to seminar

participants at the HBS-Harvard Economics Finance Lunch and participants at the HBS Research Day for

helpful comments and suggestions. We are also grateful to Martin Du§ell and Anna Christie from the UK

Debt Management O¢ ce for their help providing us with UK bond data. This material is based upon work

supported by the Harvard Business School Research Funding.
Abstract

This paper decomposes excess return predictability in ináation-indexed and nominal government bonds into liquidity, market segmentation, real interest rate risk and ináation risk.

We estimate a liquidity premium, which appears systematic in nature. It is around 40 to

70 bps during normal times but much larger during the early years of TIPS and during the

Önancial crisis in 2008-2009. We Önd evidence for large time-varying liquidity premia and

real rate risk premia in TIPS and a time-varying ináation risk premium in nominal bonds.

We Önd no evidence for segmentation between nominal and ináation-indexed bond markets

in the US or UK.
Over the 10 year period starting in 1999 the average annualized excess log return on 10

year TIPS equalled a substantial 4:16%, almost a full percentage point higher than that on

comparable nominal US government bonds. This signiÖcant return di§erential is puzzling,

because both nominal and ináation-indexed bonds are fully backed by the US government.

Moreover, the real cash áows on nominal bonds are exposed to surprise ináation while TIPS

coupons and principal are ináation protected.

This paper asks to what extent liquidity and market segmentation can explain the di§erences in yields and returns between nominal and ináation-indexed bonds. We also investigate

and quantify real cash áow risks as a source of return predictability in nominal and ináationindexed bonds and disentangle these from liquidity di§erentials.

While US nominal Treasury bonds are among the most liquid investments in the world,

TIPS empirically have a smaller and less liquid market and this was especially true during

their earlier years.2 There are reasons to believe that liquidity di§erentials between the

nominal bond market and the ináation-indexed bond market might persist beyond an initial

stage of learning and supply buildup. For any investor the riskless asset is an ináationindexed bond whose cash áows match his consumption plan (Campbell and Viceira 2001,

Wachter 2003), so that ináation-indexed bonds should typically be held by buy-and-hold

investors. Thus the return di§erential might have been the result of a liquidity discount

demanded by investors that has diminishedó but not necessarily disappearedó over time

with the growth in TIPS trading and adoption among investors.

We examine these questions adopting an empirically áexible approach. We estimate the

2For evidence of relatively lower liquidity in TIPS see Campbell, Shiller, and Viceira (2009), Gurkaynak,

Sack, and Wright (2010), Fleming and Krishnan (2009), Dudley, Roush, and Steinberg Ezer (2009).

1
liquidity di§erential between ináation-indexed bonds and nominal bonds using a variety of

proxies for liquidity. SpeciÖcally, we use the transaction volume of TIPS, the Önancing

cost for buying TIPS, the 10-year nominal o§-the-run spread and the Ginnie Mae (GNMA)

spread. These liquidity variables can explain two thirds of the time-variation in the di§erence

between US nominal and ináation-indexed bond yields.

We Önd a statistically signiÖcant and economically important time-varying liquidity component in the yield di§erential between nominal bonds and ináation-indexed bonds, popularly known among practitioners as ìbreakeven ináation.îOver our sample period the yield

on TIPS has been about 71 to 106 basis points larger on average than it would have been if

TIPS had been as liquid as nominal Treasury bonds. This high average reáects extraordinary

events associated with very low liquidity in this market. We Önd a high liquidity discount

in the years following the introduction of TIPS (about 70 to 120 bps), which we attribute to

learning and low trading volume, and during the fall of 2008 at the height of the Önancial

crisis (beyond 200 bps). We estimate a much lower but still signiÖcant liquidity discount of

between 40 to 80 bps between 2004 and 2007 and after the crisis in 2009.

We Önd that our estimated liquidity di§erential exhibits a highly signiÖcant positive

CAPM beta. This suggests that investors in TIPS bear systematic risk due to time-varying

liquidity and should be compensated for this in terms of a return premium.

We use these estimates to compute a liquidity-adjusted measure of breakeven ináation.

Breakeven ináation is often used by practitioners as a gauge of long-term ináation expectations and our Öndings suggest that liquidity-adjusted breakeven has been stable around

three percent over our sample period.

2
We next try to understand the role of liquidity and market segmentation as sources

of return-predictability in nominal and ináation-indexed bond markets. It is well-known

that the excess return on US nominal government bonds over the return on Treasury bills

exhibits predictable variation over time (Campbell and Shiller 1991, Fama and Bliss 1987,

Cochrane and Piazzesi 2005). In recent work, Páueger and Viceira (2011) provide strong

empirical evidence that the expected excess return on ináation-indexed bonds and the return

di§erential between nominal and ináation-indexed bonds are also time varying both in the

US and in the UK.

We investigate the hypothesis that the markets for nominal and ináation-indexed debt are

segmented, leading to relative price áuctuations and return predictability. Recent research

has emphasized the role of limited arbitrage and bond investorsí habitat preferences to

explain predictability in nominal bond returns (Modigliani and Sutch, 1966, Vayanos and

Vila, 2009). In the context of real versus nominal bonds, it seems plausible that the preference

of certain investorsó such as pension funds with ináation-indexed liabilitiesó for real bonds,

and the preference of othersó such as pension funds with nominal liabilitiesó for nominal

bonds might lead to imperfect market integration between both markets and this could

generate return predictability.

Following Greenwood and Vayanos (2008) and Hamilton and Wu (2010) we use the

outstanding supply of real bonds relative to total government debt as a proxy for supply

shocks in the ináation-indexed bond market. We cannot Önd any evidence for bond supply

e§ects either in the US or in the UK. One potential interpretation for this Önding could be

that governments understand investor demand for the di§erent types of securities and adjust

their issuance accordingly, e§ectively acting as an arbitrageur between the two markets.

3
Having found no evidence for market-segmentation as a source of bond return predictability, we next analyze the role of liquidity. Adjusting TIPS yields for our estimated liquidity

premium we can disentangle bond return predictability due to liquidity and due to cash

áow risks. Although government bonds in large and stable economies are generally considered default-free, their real cash áows are exposed to other risks. The prices of both

ináation-indexed and nominal government bonds change with the economy-wide real interest rate. Consequently, bond risk premia will reáect investorsí perception of real interest

rate risk, which may vary over time. The prices of nominal government bonds, but not

ináation-indexed government bonds, also vary with expected ináation, so that ináation risk

will impact their risk premia.3In this paper we Önd that liquidity is a large contributor

to return predictability in ináation-indexed bonds, but that real rate risk and ináation risk

are also statistically and economically signiÖcant contributors to return predictability in

government bonds.

This paper relates to a number of recent papers that estimate liquidity premia in TIPS.

In recent work, Fleckenstein, Longsta§ and Lustig (2010) show that ináation-swaps, which

allow investors to trade on a synthetic measure of breakeven ináation, appear to be mispriced

relative to breakeven ináation in the cash market for TIPS and nominal Treasury bonds.

In the absence of Önancing costs, synthetic breakeven ináation rates should be similar to

cash breakeven ináation but in practice they have been consistently larger. This signiÖcant

mispricing appears to be related to Önancing costs to levered investors (Campbell, Shiller,

and Viceira, 2009, and Viceira, 2011), and raises the question why unlevered investors have

not taken advantage of it. We estimate the liquidity di§erential between TIPS and Treasury

3For models of real cash áow risks in nominal government bonds see Campbell and Viceira (2001),

Campbell, Sunderam, and Viceira (2010), and Christensen, Lopez, and Rudebusch (2010).

4
bonds over a signiÖcantly longer time period and Önd it to be even higher than the pricing

di§erential in synthetic breakeven markets.

Our estimates for the liquidity premium are consistent with the magnitudes estimated

in other recent papers. DíAmico, Kim and Wei (2009) estimate a liquidity premium in a

structural model of nominal and real government bonds. Christensen and Gillan (2011)

derive a set of bounds for the liquidity premium in TIPS yields. Our approach di§ers

from other papers in that we estimate the liquidity di§erential between TIPS and nominal

Treasury bonds using a simple and transparent empirical methodology and that we focus on

the economic signiÖcance of liquidity di§erentials as a source of return predictability.

The structure of this paper is as follows. Section I estimates the liquidity premium in

US TIPS versus nominal bonds using our liquidity proxies. Section II tests the market

segmentation hypothesis in the US and in the UK, and section III considers time-varying

real interest rate risk and ináation risk premia. Finally, section IV o§ers some concluding

remarks.

I Estimating the Liquidity Component of Breakeven

Ináation

Our approach to modelling liquidity premia is empirical. We estimate the US TIPS liquidity

premium by regressing ináation compensation on measures of liquidity, following authors

such as Gurkaynak, Sack, and Wright (2010). We use four liquidity proxies: the nominal

o§-the-run spread, the GNMA spread, relative TIPS transaction volume and the di§erence

5
between TIPS asset-swap-spreads and nominal US Treasury asset-swap spreads. Since we

have data for liquidity proxies only for the US in the most recent period, this part of our

estimation is restricted to the last 10 years of US experience.

One might think that when TIPS were Örst issued in 1997, investors needed to learn

about TIPS and the market for TIPS took some time to get established. We proxy for

this idea by using the transaction volume of TIPS relative to the transaction volume of

Treasuries, a measure previously used by Gurkaynak, Sack, and Wright (2010). Fleming and

Krishnan (2009) previously found that trading activity is a good measure of cross-sectional

TIPS liquidity, lending credibility to relative transaction volume as a liquidity proxy in our

time series setup.

We interpret relative TIPS transaction volume as a measure of TIPS-speciÖc liquidity.

Following Weill (2007) and others one can also think of TIPS transaction volume as a measure

of illiquidity due to search frictions4. This should be reáected in initially low trading volumes

in TIPS and high yields during the early period.

Part of the liquidity di§erential between nominal and ináation-indexed bonds might also

be driven by ìáight-to-liquidityî episodes. In a áight to liquidity episode some market

participants suddenly prefer highly liquid securities, such as on-the-run nominal Treasury

securities, rather than less liquid securities.

The Treasury regularly issues new 10 year nominal notes, and the newest 10 year note,

also known as the "on-the-run" note or bond, is considered the most liquidly traded security

in the Treasury bond market. After the Treasury issues a new 10-year note, the prior note

4See Du¢ e, Garleanu and Pedersen (2005, 2007) and Weill (2007) for models of over-the-counter markets,

in which traders need to search for counterparties and incur opportunity or other costs while doing so.

6
goes ìo§-the-run.î The o§-the-run bond typically trades at a discount over the on-the-run

bondó i.e., it trades at a higher yieldó , despite the fact that it o§ers almost identical

cash áows with a very similar remaining time to maturity. A second type of governmentbacked bond that is also less liquidly traded than on-the-run Treasuries is GNMA bonds.

The Government National Mortgage Association (GNMA) guarantees the timely payment

of interest and principal on residential mortgage backed securities.

Flight-to-liquidity e§ects have been documented in Longsta§ (2004) for agency bonds

and Krishnamurthy (2002) for o§-the-run nominal Treasury bonds.5 We therefore interpret

the on-the-run versus o§-the-run spread in Treasury bonds and the GNMA-Treasury spreads

as empirical proxies for áight-to-liquidity episodes.

Finally, we want to capture the cost that levered investors would incur when holding

TIPS. Such investors looking for TIPS exposure can either borrow by putting the TIPS on

repo or they might consider entering into an asset swap, which requires no initial capital.

An asset swap is a derivative contract between two parties where one party receives the

cash áows on a particular government bond (e.g. TIPS or nominal) and pays LIBOR plus

the asset-swap-spread (ASW), which can be positive or negative. Our fourth measure of

liquidity is the di§erence between the asset-swap-spread for TIPS and the asset-swap-spread

for nominal Treasuries,

ASWspread

n;t = ASWT IP S

n;t  ASW$n;t (1)

This is a measure of the relative cost of Önancing a long position in the TIPS market versus

5

In the search model with partially segmented markets of Vayanos and Wang (2001) short-horizon traders

endogenously concentrate in one asset, making it more liquid. Vayanos (2004) presents a model of Önancial

intermediaries and exogenous transaction costs, where preference for liquidity is time-varying and increasing

with volatility.

7
in the nominal Treasury market. A widening of this relative spread indicates that the cost of

Önancing a long position in the TIPS market has increased relative to the cost of Önancing

a long position in the nominal Treasury market. Therefore the ASW reáects the current

and expected Önancing costs of holding the long bond position. Empirically, the asset-swapspread variable captures extraordinary events during the Önancial crisis.6

The liquidity di§erential between the nominal bond market and the ináation-indexed

bond market can also give rise to a liquidity risk premium: If the liquidity of ináationindexed bonds deteriorates during periods when investors would like to sell, as in ìáight to

liquidityî episodes, risk averse investors will demand a liquidity risk premium for holding

these bonds (Amihud, Mendelson and Pedersen 2005, Acharya and Pedersen 2005).

While the relative transaction volume of TIPS likely only captures the current ease of

trading TIPS and therefore a liquidity premium, the o§-the-run spread, the GNMA spread

and the asset-swap-spread are likely to represent both the level of liquidity and liquidity risk.

Our estimated liquidity premium is therefore likely to represent a combination of current

ease of trading TIPS versus nominal US Treasuries and the risk that the liquidity of TIPS

might deteriorate.

A Bond Notation and DeÖnitions

We denote by y

$

n;t and y

T IP S

n;t the log (or continuously compounded) yield with n periods

to maturity for nominal and ináation-indexed bonds, respectively. We use the superscript

T IP S to denote this quantity for both US and UK ináation-indexed bonds.

6See Campbell, Shiller, and Viceira (2009) for an account of liquidity events during the Fall of 2008.

8
We deÖne breakeven ináation as the di§erence between nominal and ináation-indexed

bond yields:

bn;t = y

$

n;t  y

T IP S

n;t (2)

Log excess returns on nominal and ináation-indexed zero-coupon n-period bonds held for

one period before maturity are given by

xr$

n;t+1 = ny$n;t  (n  1) y

$

n1;t+1  y

$

1;t; (3)

xrT IP S

n;t+1 = nyT IP Sn;t  (n  1) y

T IP S

n1;t+1  y

T IP S

1;t : (4)

Therefore, the log excess one-period holding return on breakeven ináation is equal to

xrb

n;t+1 = xr$n;t+1  xrT IP Sn;t+1 : (5)

Note that this is essentially the return on a portfolio long long-term nominal bonds and short

long-term ináation-indexed bonds. This portfolio will have positive returns when breakeven

ináation declines, and negative returns when it increases.

The yield spread is the di§erence between a long-term yield and a short-term yield:

s

$

n;t = y

$

n;t  y

$

1;t; (6)

s

T IP S

n;t = y

T IP S

n;t  y

T IP S

1;t ; (7)

s

b

n;t = bn;t  b1;t: (8)

Ináation-indexed bonds are commonly quoted in terms of real yields, but since xrT IP S

n;t+1

9
is an excess return over the real short rate it can be interpreted as a real or nominal excess

return. In all regressions we approximate y

$

n1;t+1 and y

T IP S

n1;t+1 with y

$

n;t+1 and y

T IP S

n;t+1 .

B Estimation Strategy

At times when TIPS are relatively less liquid than nominal bonds we would expect TIPS to

trade at a discount and the TIPS yield to increase relative to nominal yields. To account for

this premium, we estimate the following regression for breakeven ináation:

bn;t = a1 + a2Xt + "t; (9)

where Xtis a vector containing our four liquidity proxies: the o§-the-run spread, the GNMA

spread, the relative TIPS transactions volume and the di§erence between TIPS and nominal

asset-swap-spreads.

In (9) we would expect variables that indicate less liquidity in the TIPS market, such

as the o§-the-run spread, the GNMA spread and the asset-swap-spread, to enter negatively.

On the other hand higher transaction volume in the TIPS market indicates that TIPS are

easily traded and therefore it should enter positively.

The asset-swap spread reáects the Önancing costs that a levered investor incurs from

holding TIPS instead of a similar maturity nominal bonds. If the marginal investor in TIPS

is such a levered investor, we would expect breakeven ináation to fall approximately one for

one with the asset-swap-spread.

Our liquidity variables are normalized in such a way that they go to zero in a world of

10
perfect liquidity. When liquidity is perfect the o§-the-run spread, the GNMA spread and

the asset-swap spread should equal zero. The transaction volume is normalized so that its

maximum is equal to zero. That is, we assume that the liquidity premium attributable to

low transaction volume was negligible during the period of 2004-2007.

We obtain liquidity-adjusted TIPS yields by assuming that the liquidity premium estimated from the breakeven regression (9) is entirely attributable to time-varying liquidity in

TIPS rather than in nominal bonds. The estimated liquidity component in TIPS yields then

equals

L^

n;t = a^2Xt

; (10)

where a^2 is the vector of slope estimates in (9). Thus an increase in L^

n;t reáects a reduction

in the liquidity of TIPS relative to nominal Treasury bonds. Liquidity-adjusted TIPS yields

and breakeven ináation then equal

y

T IP S;adj

n;t = y

T IP S

n;t  L^

n;t; (11)

b

adj

n;t = bn;t + L^

n;t: (12)

That is, the observed yield on TIPS is larger than the liquidity-adjusted yield during times

of low liquidity and accordingly the observed breakeven ináation will be smaller than the

liquidity-adjusted breakeven ináation.

11
C Yield Data

We use data on constant-maturity ináation-indexed and nominal yields both in the US and

in the UK. Ináation-indexed bonds have been available in the UK since 1983 and in the US

since 1997. Ináation-indexed bonds are bonds whose principal adjusts automatically with the

evolution a consumer price index, which in the US is the Consumer Price Index (CPI-U) and

in the UK is the Retail Price Index (RPI). The coupons are equal to the ináation-adjusted

principal on the bond times a Öxed coupon rate. Thus the coupons on these bonds also

adjust with ináation.7

For the US we use the Gurkaynak, Sack, and Wright (2007) and Gurkaynak, Sack, and

Wright (2010, GSW henceforth) data set. GSW have constructed a zero-coupon yield curve

starting in January 1961 for nominal bonds and for TIPS starting in January 1999 by Ötting

a smoothed yield curve. The GSW data set contains constant maturity yields for maturities

of 2 to 20 years. Our empirical tests will focus on the 10-year nominal and real yields, because

this maturity bracket has the longest and most continuous history of TIPS outstanding. We

measure US ináation with the all-urban seasonally adjusted CPI, and the short-term nominal

interest rate with the 3 month T-bill rate from the Fama-Bliss riskless interest rate Öle from

CRSP. TIPS payouts are linked to the all-urban non seasonally adjusted CPI and our results

become slightly stronger when using the non seasonally adjusted CPI instead.

While the nominal principal value of TIPS increases with ináation, it is guaranteed to

never fall below its original nominal face value. As a consequence of this a recently issued

7There are further details such as in ináation lags in principal updating and tax treatment of the coupons

that slightly complicate the pricing of these bonds. More details on TIPS can be found in Viceira (2001),

Roll (2004), Campbell, Shiller, and Viceira (2009) and Gurkaynak, Sack, and Wright (2010). Campbell and

Shiller (1996) o§er a discussion of the taxation of ináation-indexed bonds.

12
TIPS, whose nominal face value is close to its original nominal face values, has a deáation

option built into it that is more valuable than that in a less recently issued TIPS with the

same time to maturity remaining. During normal times the probability of a severe and

prolonged deáation is negligible so that those bonds trade at identical prices, as discussed

in Wright (2009). Wright (2009) also points out some of the dramatic price discrepancies

between recently issued and seasoned Öve-year TIPS observed during the Önancial crisis and

attributes them to the increased value of the deáation option during the Önancial crisis. In

the Appendix we show that during the Önancial crisis our series of 10 year TIPS yields most

closely corresponds to the yields of older 10 year TIPS and conclude that it most likely does

not reáect the deáation option built into the most recent issuance of 10 year TIPS.

For the UK we use zero-coupon yield curves from the Bank of England. Anderson and

Sleath (2001) describe the spline-based techniques used to estimate the yield curves. Nominal

yields are available starting in 1970 for 0.5 to 20 years to maturity. Real yields are available

starting in 1985 for 2.5 to 25 years to maturity. We use the 20-year maturity in our tests

because 20-year nominal and real yields are available from 1985, while for instance 10-

year real yields are available only since 1991.8Ináation is measured by the non seasonally

adjusted Retail Price Index, which serves as the measure of ináation for ináation-indexed

bond payouts.

Since neither the US nor the UK governments issue ináation-indexed bills, we need to

resort to an empirical procedure to build a hypothetical short-term real interest rate. We

follow the procedure described in Páueger and Viceira (2011). For simplicity we assume that

the liquidity premium on one-quarter real bonds is zero. Finally, although our yield data

8For some months the 20 year yields are not available and instead we use the longest maturity available.

The maturity used for the 20 year yield series drops down to 16.5 years for a short period in 1991.

13
sets are available at a monthly frequency, we focus on quarterly overlapping bond returns

in order to reduce the ináuence of high-frequency noise in observed ináation and short-term

nominal interest rate volatility in our tests.

D Data on Liquidity Proxies

We obtain the 10 year o§-the-run spread from the Federal Reserve and from Bloomberg.9

While GNMA bonds do not contain any default risk, they do contain prepayment risk,

because mortgage holders can prepay without penalty. We obtain a GNMA spread adjusted

for prepayment risk from Bloomberg.10

We obtain Primary Dealersítransaction volumes for TIPS and nominal Treasury securities from the New York Federal Reserve FR-2004 survey. We construct our measure of

relative transaction volume as log T ransT IP S

t

=T rans$

t



, where T ransT IP S

t denotes the average weekly transactions volume over the past month and T rans$

t

the corresponding Ögure

for nominal bonds. For T rans$ we use the transaction volume of government coupon securities with at least 6 (before 2001) or 7 (from 2001) years to maturity. We choose the

transaction volume series for coupon bonds with a long time to maturity because we are

aiming at capturing the di§erential liquidity of TIPS with respect to 10 year nominal bonds.

Including all maturities or even T-bills would also reáect liquidity of short-term instruments

versus long-term instruments. We then smooth the measure of relative transaction volume

over the past three months because we think of it as capturing secular learning e§ects rather

9The on the run data is from Bloomberg (USGG10YR), and the o§ the run is from the Federal Reserve

publication H.15 ìInterest Ratesî.

10Ticker GNSF060. This is the prepayment-option adjusted spread based on a 6% coupon 30 year GNMA

generic bond. It is adjusted for prepayment risk using the Bloomberg prepayment model.

14
than capturing short-term áuctuations in liquidity.11 We normalize the relative transaction

volume so that its maximal value is equal to zero.

We obtain asset-swap-spread data from Barclays Live. We only have data on ASWspread

n;t

from July 2007 until April 2009, and set it to its July 2007 value of 40 bps when the assetswap-spread series is not available. For the 10 year TIPS asset-swap-spread we use the

July 2017 Asset Swap and for the 10 year nominal Asset Swap we use the generic 10 Year

On-the-Run Par asset-swap-spread.

Figure 1 shows our four liquidity variables. The dissimilar time-series patterns of the

variables suggest that each one represents a di§erent aspect of market liquidity, although

the spread variables all jump during the Önancial crisis of 2008-2009. The on-the-run o§-therun spread exhibits high frequency variation. The GNMA spread, on the other hand, moves

relatively slowly. One reason for the di§erence in the two spreads could be that they have a

di§erent investor base. The GNMA spread pattern of a lower spread between 2002 and 2007

agrees with anecdotes of long-term investors who were particularly willing to invest into less

liquid securities in order gain yield during that period. The relative transaction volume rises

linearly through 2004 and then stabilizes. This suggests that the liquidity premium due to

the novelty of TIPS should have been modest in the period since 2004.

Finally the asset-swap-spread variable ASWspread

n;t varies within a relatively narrow range

of 35 basis point to 41 basis points from July 2007 through August 2008, and it rises sharply

during the Önancial crisis, reaching 130 bps in December 2008. That is, before the crisis

11 In 2001 the Federal Reserve changed the maturity cuto§s for which the transaction volumes are reported.

Before 6/28/2001 we use the transaction volume of Treasuries with 6 or more years to maturity while starting

6/28/2001 we use the transaction volume of Treasuries with 7 or more years to maturity. The series after

the break is scaled so that the growth in T rans$from 6/21/2001 to 6/28/2001 is equal to the growth in

transaction volume of all government coupon securities.

15
Önancing a long position in TIPS was about 40 basis more expensive than Önancing a long

position in nominal Treasury bonds, but this cost di§erential rose to more than 120 basis

points after the Lehman bankruptcy in September 2008. Campbell, Shiller, and Viceira

(2009) argue that the bankruptcy of Lehman Brothers in September of 2008 had a signiÖcant

e§ect on liquidity in the TIPS market, because Lehman Brothers had been very active in

the TIPS market. The unwinding of its large TIPS inventory in the weeks following its

bankruptcy, combined with a sudden increase in the cost of Önancing long positions in TIPS

appears to have induced an unexpected downward price pressure in the TIPS market. This

led to a liquidity-induced sharp tightening of breakeven ináation associated with a widening

of the TIPS asset-swap-spread.

Table I shows summary statistics for bond spreads and excess returns as well as liquidity

variables. One can see that over our sample period breakeven ináation has averaged 2:25%

and TIPS yields have averaged 2:55%. The realized average annualized log excess returns

on TIPS have been equal to 4:16%, exceeding the log excess returns on nominal government

bonds by 91 basis points (bps). The average log excess returns on UK ináation-indexed

bonds, on the other hand, have been substantially smaller at only 1:66% over the longer

sample period from 1985.4 to 2009.12.

E Estimation Results

Table II reports OLS estimates of the regression (9), adding the liquidity proxies one at a

time. Columns 1 through 3 always include ASWspread

n;t , but with a slope set to its theoretical

value of 1. Column 4 presents estimates with freely estimated coe¢ cients for all four

liquidity proxies.

16
Table II shows coe¢ cients whose signs are consistent with expectations and generally

statistically signiÖcant. Interestingly, our liquidity measures explain a very large fraction

of the variability of breakeven ináation, from 44% in column 1 to 67% in column 4. The

R2increases with every additional liquidity control introduced, indicating that each of the

controls helps explain the liquidity premium on TIPS.

The negative and signiÖcant coe¢ cient on relative TIPS trading volume in Table II

indicates that the impact of search frictions on ináation-indexed bond prices appears to have

been exacerbated during the early period of ináation-indexed bond issuance, when maybe

only a small number of sophisticated investors had a good understanding of the mechanics

and pricing of these new bonds.

In Table II breakeven ináation moves negatively with both the on-the-run versus o§-therun spread in Treasury bonds and the GNMA-Treasury spreads in our sample period. This

indicates that áight-to-liquidity episodes and more generally investorsípreference for trading

on-the-run nominal Treasury bonds also help explain the yield di§erential between nominal

yields and ináation-indexed yields. This empirical Önding indicates that while during a

áight-to-liquidity episode investors rush into nominal on-the-run US Treasuries, they do not

buy US TIPS to the same degree. This is especially interesting given that both types of

bonds are fully backed by the same issuer, the US Treasury.

Column 4 in Table II shows that the freely estimated coe¢ cient on the asset-swap-spread

di§erential is at 1:59 somewhat larger in absolute value than 1. During the Önancial crisis

securities markets were severely disrupted and the buyers and sellers of asset swaps may not

have acted as the marginal buyers and sellers of TIPS. Estimating freely the regression

coe¢ cient on the asset-swap-spread accounts for the possibility that the asset-swap-spread

17
only represents a fraction of the Önancing cost for the marginal holder of TIPS. The standard

error on the regression coe¢ cient indicates that it is precisely estimated. The large size of

this parameter suggests the relevance of liquidity factors in explaining the sharp fall in

breakeven during the Önancial crisis, since the asset-swap-spread di§erential behaves almost

like a dummy variable that spikes up during the Önancial crisis. Nonetheless, the di§erence

between the liquidity component estimated in columns 3 and 4 appears small as indicated

by the very similar R2 and we will focus on the freely estimated version from column 4 for

its áexibility.

One might be concerned that our estimation of the liquidity premium L relies on extrapolation outside the range of historically observed liquidity events. The e§ect of our liquidity

proxies on the liquidity premium in TIPS might be nonlinear and this might lead to signiÖcant errors in the extreme values of the estimated liquidity premium. We therefore include

additional quadratic terms into the estimation of the liquidity premium in column 5. Further

results including interaction terms are reported in the Appendix. The squared o§-the-run

spread and GNMA spread enter signiÖcantly and while their addition only yields a small

improvement in the R2from 67% to 70% one could imagine that their impact would be substantial on our liquidity estimates during events of extreme liquidity or extreme illiquidity.

The results are not sensitive to the inclusion of the Önancial crisis in the sample period.

In column 6 we obtain very similar regression coe¢ cients and an R2 of 47% over a sample

period from 1999 to 2006.

If ináation expectations are correlated with liquidity in the TIPS market one might be

worried that this a§ects our estimation of the liquidity premium in Table II. We therefore

complement our regressions with survey ináation expectations as in DíAmico, Kim and Wei

18
(2008). In order to control for movements in expected ináation we consider the di§erence

between breakeven ináation and survey ináation expectations and regress this di§erence

onto our liquidity proxies. We use the 10 year forecast of CPI ináation from the Survey of

Professional forecasters, which is available at a quarterly frequency and is released towards

the end of the middle month of the quarter. We therefore can only use quarterly data for this

exercise and we use breakeven ináation and our liquidity proxies at the end of the second

month of each quarter.

Table II, columns 7 and 8 show the quarterly regression of breakeven ináation and of the

di§erence between breakeven ináation and survey expectations onto the liquidity proxies.

Newey-West standard errors with four lags are shown in brackets. The regression coe¢ cients

are similar to the monthly regressions reported in the other columns of Table II. If anything

the signiÖcance of the coe¢ cients increases when taking out survey expectations.

Figure 2 shows our liquidity premium as estimated in our unconstrained linear speciÖcation in column 4 as well as the liquidity premium as estimated in columns 3 and 5. We will

refer to column 3 as the "constrained" estimate, because the coe¢ cient on the asset-swapspread is set to 1: The estimate from column 5 with quadratic terms will be referred to as

"nonlinear".

Our unconstrained linear speciÖcation yields an average spread due to liquidity of around

106 bps. Although this average is high, one must take into account that it reáects periods of

very low liquidity in this market. Figure 2 shows a high liquidity premium in the early 2000ís

(about 120 bps), but a much lower liquidity premium between 2004 and 2007 (70 bps). The

premium shoots up again beyond 200 bps during the crisis, and Önally comes down to 70

bps after the crisis.

19
Our alternate speciÖcations yield somewhat lower average values for liquidity premium

at 71 bps (nonlinear estimate) and 89 bps (constrained estimate) but they show extremely

similar time series variation. The nonlinear speciÖcation suggests that the liquidity premium

in normal times might be as low as 40 bps and the liquidity premium during the initial period

of TIPS might have been around 70 to 80 bps. However, all three speciÖcations suggest that

the liquidity premium reached prior unprecedented levels of over 200 bps during the Önancial

crisis.

The time series of our liquidity premium is consistent with the Öndings in DíAmico,

Kim and Wei (2008) but the level of the unconstrained linear estimate is a little higher. In

recent work Fleckenstein, Longsta§ and Lustig (2010) show that ináation swaps, which allow

investors to trade on ináation without putting up any initial capital, appear to be mispriced

relative to breakeven ináation in the cash market for TIPS and nominal Treasury bonds.

Their series of average mispricing between synthetic and cash breakeven ináation resembles

our series of di§erential Önancing costs ASWspread

n;t both in terms of level and time series

variation. Our estimated liquidity premium could be interpreted as a measure of mispricing

of TIPS relative to nominal Treasury bonds and we Önd this to be even larger than the

mispricing relative to synthetic instruments.

The high liquidity premium in TIPS is puzzling given that bid-ask spreads on TIPS are

small. As previously noted by Wright (2009) it seems implausible that the liquidity premium

in TIPS yields simply serves to amortize transaction costs of a long-term investor. Haubrich,

Pennachi and Ritchken (2010) report bid-ask spreads for TIPS between 0.5 bps up to a high

of 10 bps during the Önancial crisis. As argued before, TIPS should typically be held by

buy-and-hold investors. In a simple model of liquidity, such as given in Amihud, Mendelson

20
and Pedersen (2005), a transaction cost of 10 bps can only justify a liquidity premium of

1bp for a 10 year TIPS that is held by a buy-and-hold investor.

It is instructive to compare the liquidity premium for TIPS to the liquidity premium for

o§-the-run nominal Treasuries, adjusted for the likely time to convergence of the liquidity

di§erential. The average o§-the-run spread over our sample period is about 4 bps. However,

the on-the-run o§-the-run liquidity di§erential can be expected to converge when the new

on-the-run nominal 10 year bond is issued. Therefore, convergence of the on-the-run o§-

the-run liquidity di§erential should take at most six months. Thus the annualized return

on the liquidity di§erential between 10 year on-the-run and o§-the-run nominal Treasury

bonds is about 80 bps. On the other hand, the TIPS liquidity premium might only converge

throughout the life of the bond and hence persist for 10 years for a 10 year TIPS. The

annualized return on liquidity from 10-year zero-coupon TIPS then should be approximately

equal to our estimate of the liquidity premium, i.e., between 40bps and 70bps during normal

times. These simple calculations imply that the estimated liquidity premium in TIPS, though

puzzlingly large when compared to bid-ask spreads, give rise to liquidity returns which are

comparable in magnitude to those on o§-the-run nominal Treasuries.

Figures 3 and 4 show liquidity-adjusted breakeven ináation and TIPS yields, respectively.

The regression constant in Table II can be interpreted as the average liquidity-adjusted

breakeven ináation for the respective speciÖcation of the liquidity premium. While the unconstrained estimation suggests an average liquidity-adjusted breakeven over our sample

period of 3:3%, the nonlinear estimate suggests an average breakeven ináation of only 3:0%:

We can also see that liquidity-adjusted breakeven ináation is much more stable than nonadjusted breakeven ináation. Table II , column 4 shows an R2 of 67%; implying that the

21
variance of liquidity-adjusted breakeven is only one third of the variance of non-adjusted

breakeven. Both speciÖcations indicate that liquidity-adjusted breakeven ináation has been

stable over our sample period and attribute the drop in breakeven ináation during the fall

of 2008 to liquidity. Figure 4 shows that if TIPS had remained as liquid as nominal Treasuries their yields would have dropped dramatically in the fall of 2008. This has important

implications for the interpretation of the dramatic reduction in breakeven ináation observed

during the Önancial crisis as an indicator of massive expected deáation among bond market

participants.

II Testing For Market Segmentation E§ects

Páueger and Viceira (2011) documented economically and statistically signiÖcant excess

return predictability in ináation-indexed bonds and in the di§erence between nominal bonds

and ináation-indexed bonds. In this section we explore whether shocks to the relative supply

of nominal and ináation-indexed bonds can explain these Öndings.

The preferred-habitat hypothesis of Modigliani and Sutch (1966) states that the preference of certain types of investors for speciÖc bond maturities might result in supply imbalances and price pressure in the bond market. In recent work Vayanos and Vila (2009)

formalize this hypothesis in a theory where risk averse arbitrageurs do not fully o§set the

price imbalances generated by the presence of preferred-habitat investors in the bond market,

leading to excess bond return predictability. Greenwood and Vayanos (2008) and Hamilton

and Wu (2010) Önd empirical support for this theory using the relative supply of nominal

Treasury bonds at di§erent maturities as a proxy for supply shocks.

22
Following Greenwood and Vayanos (2008) we try to control for the potential segmentation

between both markets and supply e§ects using the outstanding supply of real bonds relative

to total government debt. If supply is subject to exogenous shocks while clientele demand

is stable over time we would expect increases in the relative supply of ináation-indexed

bonds to be correlated with contemporary decreases in breakeven ináation, as the price of

ináation-indexed bonds falls in response to excess supply. Subsequently we would expect to

see positive returns on ináation-indexed bonds as their prices rebound.

Alternatively, it could be the case that bond demand changes over time, and the government tries to accommodate changes in demand. This would be consistent with a debt

management policy that tries to take advantage of interest rate di§erentials across both

markets. In this case we would expect the relative supply of ináation-indexed bonds to be

unrelated to subsequent returns, and possibly to be even positively correlated with contemporaneous breakeven ináation.

We measure the relative supply of ináation-indexed bonds in the US as the nominal

amount of TIPS outstanding relative to US government TIPS, notes and bonds outstanding.12 The relative supply variable for the UK is computed similarly, as the total amount of

ináation-linked gilts relative to the total amount of conventional gilts outstanding. 13

12The economic report of the president reports US Treasury securities by kind of obligation and reports

T-bills, Treasury notes, Treasury bonds and TIPS separately. The data can be found in Table 85 for the

reports until 2000 and in Table 87 in subsequent reports at http://www.gpoaccess.gov/eop/download.html.

The face value of TIPS outstanding available in the data is the original face value at issuance times the

ináation incurred since then and therefore it increases with ináation. The numbers include both privately

held Treasury securities and Federal Reserve and intragovernmental holdings as in Greenwood and Vayanos

(2008).

13We are deeply grateful to the UK Debt Management O¢ ce for providing us with the UK data. Conventional gilts exclude áoating-rate and double-dated gilts but include undated gilts. The face value of

index-linked gilts does not include ináation-uplift and is reported as the original nominal issuance value.

Our results are not sensitive to including or excluding the ináation uplift.

23
Let DT IP S

t denote the face value of ináation-indexed bonds outstanding and Dt the

combined face value of nominal and ináation-indexed bonds outstanding at time t for either

the US or the UK. We deÖne the relative supply as

Supplyt = D

T IP S

t

=Dt: (13)

We also consider the relative issuance Supplyt, which we compute as

Supplyt =



D

T IP S

t  D

T IP S

t1



=DT IP S

t1  (Dt  Dt1) =Dt1: (14)

Figure 5A plots the relative supply of TIPS, DT IP S

t

=Dt, and 10 year breakeven ináation

in the US. It illustrates the rapid increase in the relative amount of TIPS outstanding.

Starting from less than 2% in 1997 TIPS increased to represent over 14% of the US Treasury

coupon bond portfolio in 2008. Subsequently to the Önancial crisis the US government issued

substantial amounts of nominal notes and bonds, leading to a drop in the relative TIPS share

in 2009. At the same time the level of breakeven ináation remained relatively steady over

this 11 year period with a large drop in the fall of 2008, as discussed earlier.

Figure 5B illustrates the history of the relative share of UK ináation-linked gilts outstanding. The relative share of linkers has increased over the period from about 8% in 1985

to over 17% in 2008. At the same time 20 year UK breakeven ináation has fallen in the

period 1985-2009, reaching a low of 2.1% in 1998. The increase in ináation-linked bonds

outstanding accelerated noticeably after 2004. Greenwood and Vayanos (2009) analyze this

episode in light of the UK Pensions Act of 2004, which provided pension funds with a strong

incentive to buy long-maturity and ináation-linked government bonds and subsequently led

24
the government to increase issuance of long-maturity and ináation-linked bonds.

Table II shows regressions of breakeven ináation onto the relative supply of ináationindexed bonds. If markets are segmented and subject to exogenous supply shocks we would

expect a negative coe¢ cient on measures of relative supply of ináation-indexed bonds. In

the US neither the relative supply nor relative issuance Supplyt appear to be related to

breakeven ináation. The results do not change when we add our liquidity proxies or a time

trend as controls.

One might be concerned that the relative supply of TIPS has been very predictable and

therefore does not represent an adequate measure of supply shocks. In the Appendix we

conduct Dickey-Fuller tests to Önd that in the US we cannot reject a unit root in Supplyt or

in Supplyt: However, the year-over-year change in relative issuance appears stationary and

we construct a supply shock "

Supply

t as the residual from an autoregression of Supplyt 

Supplyt12 with twelve lags. In the UK we can reject stationarity in relative issuance

Supplyt; potentially reáecting the less regular issuance cycle in the UK, and we construct

a supply shock "

Supply

t as the residual from an autoregression of Supplyt with twelve lags.

We add the residual shock "

Supply

t as a potential explanatory variable to our regressions in

Table II.

The US supply shock "

Supplyt

t by itself covaries positively with breakeven ináation and is

marginally signiÖcant but this e§ect goes away when we add the liquidity controls. Those

results suggest that the US government might react to yield di§erentials and moreover might

react to di§erential liquidity between the nominal and ináation-indexed markets.

Table IIB shows regressions of UK breakeven ináation onto the relative supply. Due to

25
data constraints we are not able to control for liquidity. The results are very similar to the

US results, even though the maturities of the bonds and the sample periods are di§erent:

The supply variable is signiÖcant but it switches sign as we include a time trend in the

regression, while the change in supply does not enter signiÖcantly. The time trend is highly

statistically signiÖcant and increases the R2from 26% to 65%. The supply shock variable

"

Supply

t does not appear to commove with breakeven ináation either positively or negatively.

Figure 5B helps to interpret the supply coe¢ cientís sign change when adding a time trend.

Since the mid-1980ís the supply of ináation linkers in the UK has risen, while breakeven ináation has been generally declining. This secular decline in breakeven ináation likely reáects

changes in monetary policy and declines in both realized and expected ináation (Campbell,

Shiller, and Viceira 2009), rather than changes in bond supply. A simple regression of UK

breakeven on the supply of ináation linkers therefore gives a negative slope. Introducing a

time trend takes care of this common inverse trend, and switches the sign of the slope on

the supply variable to positive. This positive partial correlation might again lend weight to

the interpretation that the UK government reacts to increased demand for ináation-linked

bonds by issuing more ináation-indexed bonds, an interpretation that would be consistent

with the episode described in Greenwood and Vayanos (2009).

If markets are segmented we would expect a positive shock in the relative supply of

ináation-indexed bonds to predict lower excess returns on nominal bonds over ináationindexed bonds. Our left-hand-side variables in Table IV are the nominal, ináation-indexed

and breakeven returns as deÖned in (3), (4) and (5). Páueger and Viceira (2011) show that

nominal, TIPS and breakeven term spreads are signiÖcant predictors of the corresponding

excess returns and therefore we control for these spreads in our regressions. One might also

26
think that liquidity should predict bond excess returns and we therefore include it as an

additional control.

Panel A in Table IV shows that in the US the supply variables do not help to predict

bond excess returns. TIPS and breakeven term spreads still enter signiÖcantly and predict

TIPS excess returns and breakeven excess returns, respectively, after including liquidity and

supply e§ects. The liquidity premium enters signiÖcantly and in particular helps predict the

breakeven return. The results are extremely similar for the UK.

Overall, we Önd no evidence of supply shocks predicting bond excess returns. In particular

our measures of relative supply shocks cannot explain why term spreads predict excess returns

on ináation-indexed bonds and on nominal bonds in excess of ináation-indexed bonds as

found in Páueger and Viceira (2011).

In summary, there is no evidence of relative supply shocks predicting bond excess returns

in either the UK or the US. We Önd some weak evidence that relative supply of ináationindexed bonds moves positively with breakeven ináation. These results do not seem consistent with segmented markets that are subject to exogenous supply shocks. Instead they

might be consistent with the US and UK governments accommodating demand pressures

from investors for nominal or ináation-indexed bonds. Panel A in Table III shows that liquidity is also a very strong predictor of breakeven excess returns in the US. We therefore

proceed to analyze the role of liquidity more closely by decomposing breakeven ináation into

a liquidity component and liquidity-adjusted breakeven ináation.

27
III Time-Variation of Real Interest Rate and Ináation

Risk Premia

A Predictive regressions with liquidity-adjusted yields and returns

In this section we decompose return-predictability in nominal and ináation-indexed bonds.

Ináation-indexed bond return predictability could be the result of either a time-varying real

interest rate risk premium, a time-varying liquidity premium, or a combination of bothó

since supply e§ects do not seem to matter. Breakeven and nominal bond excess return

predictability could be the result of a time-varying ináation-risk premium and time-varying

liquidity.

We can use our estimates of liquidity e§ects on ináation-indexed bond prices and returns

to disentangle these e§ects. We start by replacing the TIPS yield by the liquidity-adjusted

TIPS yield (11) and breakeven by liquidity-adjusted breakeven (12) and use these to compute

liquidity-adjusted TIPS and breakeven returns according to

xrT IP SL

n;t+1 = ny

T IP S;adj

n;t  (n  1) y

T IP S;adj

n1;t+1  y

T IP S

1;t ; (15)

xrb+L

n;t+1 = xr$n;t+1  xrT IP SLn;t+1 : (16)

We also examine whether there is evidence of a time-varying liquidity risk premium, by

28
looking at the predictability of the liquidity return. We deÖne the liquidity return as

r

L

n;t+1 =  (n  1) Ln1;t+1 + nLn;t: (17)

We can think of r

L

n;t+1 as the return on TIPS return due to time-varying liquidity.

Our estimates for the liquidity premium Ln;t are based on the full-period regression

allowing for a áexible regression coe¢ cient on the asset-swap spread, reported in Table II,

column 4 but our results are not sensitive to the particular speciÖcation of Ln;t; as we show

in the Appendix.

Table V shows that conditional on our estimates of liquidity-adjusted yields and returns,

the real yield spread forecasts positively ináation-indexed bond returns, and the breakeven

ináation spread forecasts breakeven returnsó or the return on nominal bonds in excess of the

return on ináation-indexed bonds. This provides support for the hypothesis that real and

nominal bond term spreads reáect time-varying real interest rate and ináation risk premia.

Remarkably the liquidity variable does not predict liquidity-adjusted real bond excess returns

or breakeven excess returns. The current level of the liquidity premium does not appear to

be related to fundamental cash-áow risk as represented by the real interest rate risk premium

or the ináation risk premium.

The results shown use excess returns on TIPS over our hypothetical real short rate.

However, the return-predictability results in Table V hold up if instead we consider nominal

returns on TIPS in excess of the nominal short rate, as shown in the Appendix.

The last column of Table V reports a regression of the liquidity return r

L

n;t+1 onto the

liquidity-adjusted real term premium, the liquidity-adjusted breakeven spread, and Ln;t. We

29
Önd that the liquidity return is predictable from the liquidity premium with a large and highly

signiÖcant regression coe¢ cient, suggesting that the return-predictability in TIPS found in

Páueger and Viceira (2011) is at least partly due to a time-varying and predictable liquidity

premium. The e§ect of the liquidity premium on returns is such that when liquidity in the

TIPS market is scarce, TIPS enjoy a higher expected return relatively to nominal bonds,

rewarding investors who are willing to invest into a temporarily less liquid market.

The results shown in Table V strongly suggest that return-predictability in ináationindexed and nominal government bonds in Páueger and Viceira (2011) is driven by a combination of factors. Our results o§er support for a time-varying real interest rate risk premium

and a time-varying liquidity premium in TIPS and for a time-varying ináation risk premium

in nominal bonds.

B Historical Fitted Risk Premia and Systematic Risk

We next aim to better understand the economic signiÖcance of the di§erent components of

bond returns. SpeciÖcally, Table VI compares the market-loadings and predictability of the

di§erent bond return components. 14

Our estimates attribute the negative average breakeven ináation returns over our sample

period to the large liquidity e§ects in TIPS.15 After adjusting for liquidity, we obtain a

14 In Panel A we obtain liquidity-adjusted expected excess log returns and expected liquidity returns as

Ötted values from the regressions shown in columns 1, 3 and 4 in Table V. Our average return calculations

are based on log returns with no variance adjustments for Jensenís inequality. We compute CAPM betas

using the stock market as the proxy for the wealth portfolio. The US excess stock return is the log quarterly

return on the value-weighted CRSP index, rebalanced annually, in excess of our log 3 month interest rate.

The UK excess stock return similarly is computed as the log quarterly total return on the FTSE in excess

of our log 3 month interest rate.

15 In Panel A, the moments of non liquidity-adjusted excess log returns are for a slightly longer sample pe30
positive breakeven ináation return of 74 bps p.a. on average. The estimated average TIPS

return due to time-variation in liquidity is large at 1.36% p.a.. At the same time our estimates

imply that even controlling for liquidity the average return on TIPS has been positive at

2.87% p.a..

The second column in the panel investigates the extent to which excess returns are

systematic. The upper half of the panel shows that over our sample period the CAPM beta

of nominal bonds and breakeven ináation have been signiÖcantly negative, both economically

and statistically, while the CAPM beta of TIPS has been essentially zero. The lower half of

the panel shows that by contrast the CAPM beta on liquidity-adjusted breakeven ináation

returns has been zero, and the CAPM beta liquidity-adjusted TIPS returns has been negative

and highly signiÖcant at 0:46. We also estimate a positive and highly signiÖcant beta of

0:53 on the returns on the liquidity premium.16

Our estimates of CAPM betas of liquidity-adjusted returns suggest that liquidity risk

explains most of the systematic variation in breakeven ináation returns. The positive beta

on liquidity implies that TIPS tend to become illiquid relative to nominal Treasury bondsó

or conversely, Treasury bonds become more liquid relative to TIPSó at times when the stock

valuations fall. The strong positive covariation of our liquidity estimate with stock returns

suggests that investors should expect to earn a premium on TIPS for bearing systematic

variation in liquidity.

riod (1999.4-2009.12) than the moments of liquidity-adjusted returns and liquidity returns (1999.6-2009.12).

This explains why the average liquidity-adjusted excess log return on TIPS and the log return on liquidity

do not exactly add up to the average excess log return on TIPS.

16Our results are robust to excluding the Önancial crisis period from the estimation of CAPM betas.

Excess log returns on TIPS, nominal Treasuries, breakeven ináation, and our measure of liquidity do not

load signiÖcantly on Fama-French value and size factors. For details see the Appendix.

31
Nevertheless, systematic risk does not fully explain the returns on liquidity over our

sample period. In the Appendix we show that after controlling for market returns the

average return on TIPS liquidity is still positive. The market alpha of liquidity returns is

statistically signiÖcant for the full time period but not for a more recent sub-period starting in

2002. Adding additional controls for size, value, and the Pastor-Stambaugh (2003) liquidity

factor also renders the intercept insigniÖcant. These results suggests that the positive alpha

was at least partly related to the novelty of TIPS and due to particular market conditions

during the Örst few years of TIPS issuance. Once TIPS were more solidly established the

intercept was still positive but no longer statistically signiÖcant and it seems plausible to

think that this positive intercept will be smaller going forward.

The negative beta of liquidity-adjusted TIPS returns imply a pro-cyclical behavior of

real interest rates in the framework of Campbell, Sunderam, and Viceira (2011). Once we

adjust for liquidity we Önd a breakeven ináation beta that is close to zero, so that systematic

ináation risk appears to have been negligible during our sample period.17 Procyclical behavior of real interest rates and low ináation risk makes all Treasury bonds, whether nominal

or ináation-indexed, safe assets that provide investors with sizable diversiÖcation beneÖts

relative to stocks.

In order to understand the economic signiÖcance of the estimated return predictability,

we now calculate the volatilities of predicted returns and compare them to the variability of

realized returns. We obtain the expected liquidity-adjusted breakeven returns, the expected

liquidity-adjusted TIPS returns and expected liquidity returns from the return-predictive

17These results suggest that the negative ináation risk premium estimated by Campbell, Sunderam and

Viceira (2010) over our sample period might have been partly caused by systematic variation in the relative

liquidity of TIPS and nominal Treasury bonds.

32
regressions in Table V. We obtain the expected excess returns on nominal bonds, TIPS

and breakeven ináation as in Páueger and Viceira (2011). They specify the nominal risk

premium at any point in time as the expected excess log return on nominal bonds predicted

by the nominal term spread. They obtain TIPS and breakeven risk premia as Ötted values

of regressions analogous to those in Table V, column 1 and Table V, column 3 without any

liquidity adjustments.

From the second column of Panel A in Table VI we see that expected liquidity returns

were the most volatile component of predictable returns. Time-varying expected liquidity

returns can account for 12% of the variance of realized TIPS returns, while time-variation in

the expected returns on liquidity-adjusted TIPS explain 6% of the variance of realized TIPS

returns. The time-variation in expected returns on liquidity-adjusted breakeven ináation

and TIPS can account for 3% and 5% of the sample variability of realized nominal bond

returns, respectively.

For the alternate estimates of the liquidity premium we obtain a lower but positive average

liquidity returns and a lower but positive average liquidity-adjusted breakeven returns but

betas and volatilities look similar. More details can be found in the Appendix.

Figure 6A illustrates the time series of the Ötted US expected excess returns. We interpret

the expected liquidity excess return as a liquidity return premium, the expected liquidityadjusted breakeven return as an ináation risk premium and expected liquidity-adjusted TIPS

returns as a real rate risk premium. It shows that during the period of 2000 to 2006 the

ináation risk premium was small or negative. During the period of high oil prices in 2008

and during the peak of the Önancial crisis in late 2008 the ináation risk premium was positive

but subsequently fell to almost -10% at the end of 2009, precisely at a time when the real

33
rate risk premium increased sharply. The liquidity risk premium on TIPS was large in the

early 2000ís, but declined steadily during the decade, with the exception of a pronounced

spike during the Önancial crisis in the Fall of 2008.

Panel B in Table VI shows similar statistics for UK excess bond returns over the longer

sample period 1985-2009. Due to data constraints we are not able to compute a liquidityadjustment for the UK. However, arguably liquidity-adjustments in the UK bond market are

likely to be less signiÖcant than in the US bond market. UK ináation-linked bonds have been

issued for a signiÖcantly longer period and therefore it appears plausible that initial learning

should a§ect only a small portion of their time series. Moreover, neither UK nominal nor

ináation-indexed bonds are likely to enjoy the same extraordinary liquidity beneÖts as US

nominal Treasury bonds, suggesting that ìáight-to-liquidityîe§ects should be less signiÖcant

in generating a liquidity di§erential between ináation-indexed and nominal bonds in the UK.

We Önd an average excess return on ináation-indexed bonds of 1.66% p.a., and an average

excess return on ináation-indexed bonds of 1.81% per annum in UK bonds. If instead

UK ináation-indexed bonds were less liquid than nominal UK bonds then the liquidityadjusted breakeven excess return might be slightly higher and the excess return on ináationindexed bonds might be slightly lower. Both of these components of UK bond risk premia

appear economically signiÖcant. In particular, we estimate a slightly larger excess return on

breakeven ináation for UK bonds than for US bonds.

The CAPM beta of UK ináation-indexed bonds at 0:15 is positive and statistically signiÖcant, while the CAPM beta of breakeven ináation is essentially zero over the full sample.

The betas on UK bonds indicate that on average ináation-indexed bonds, and to some degree nominal bonds, have been risky over the time period under consideration. On the other

34
hand, breakeven ináation has not exhibited any systematic risk. However, the UK betas

hide important variation across sub-periods. In the Appendix we show that the beta of

UK breakeven ináation was positive in the period prior to 1999 and turned negative and

signiÖcant in the post-1998 period, suggesting that ináation risk in nominal bonds was much

more pronounced in the Örst half of the sample.

Interestingly, Figure 6B suggests that the UK breakeven risk premium shot up during the

Önancial crisis and in contrast to the US has remained high. In the framework of Campbell,

Sunderam, and Viceira (2010) this could indicate that while investors in the UK fear that

further economic deterioration will go along with ináation, US investors are concerned about

low growth accompanied by low ináation or even deáation.

IV Conclusion

This paper explores the sources, magnitude, and time variation in bond risk premia in US

and UK ináation-indexed and nominal bonds. We Önd strong empirical evidence for two

di§erent potential sources of excess return predictability in ináation-indexed bonds: real

interest rate risk and liquidity. We also provide empirical evidence that nominal bond return

predictability is related to both time variation in the real interest rate risk premium and time

variation in the ináation risk premium.

Using a variety of liquidity indicators, we Önd that the liquidity premium on TIPS yields

relative to nominal Treasury bond yields exhibits strong time-variation, with a large premium

in the vicinity of 70 to 120 bps early in the life of TIPS, a signiÖcant decline to a lower

premium of 40 to 70 bps after 2004, and a sharp increase to over 200 bps during the height

35
of the Önancial crisis in the fall of 2008 and winter of 2009. Since then, the premium has

declined back to 50 to 80 bps.

Time-varying liquidity explains up to 67% of the time-series variance of breakeven ináation. Once we adjust breakeven ináation for liquidity e§ects, we Önd it to be stable over our

sample period around 3%, suggesting that bond investorsílong-term ináation expectations

in the US have not moved signiÖcantly, even during the Önancial crisis.

In our analysis of price pressures due to supply shocks in the ináation-indexed bond

market we Önd no evidence for a supply channel in either the US or in the UK. If anything,

our results are consistent with the government trying to accommodate shifts in the demand

for nominal bonds, relative to ináation-indexed bonds.

A high liquidity return can explain why US TIPS have exhibited higher excess returns

than nominal Treasuries over the 1999-2009 period. Returns due to liquidity appear to be

highly systematic and are correlated with aggregate stock returns, indicating that investors

should expect to be paid a return premium for bearing this risk.

Variation in expected liquidity excess returns explains a substantial 12% of the observed

variation in TIPS returns. The liquidity premium does not predict liquidity-adjusted returns

on TIPS or on breakeven, so that it does not seem to proxy for any real interest rate risk or

ináation risk. Hence, while we Önd that the liquidity di§erential between nominal bonds and

TIPS moves with the aggregate market, it bears no relationship to the real cash áow risks

of the speciÖc securities, a conclusion that should be important in guiding future models of

liquidity.

Our liquidity premium is correlated with other liquidity spreads in Öxed income markets,

36
particularly the o§-the-run spread (Krishnamurthy, 2002), and it exhibits an economically

and statistically signiÖcant positive CAPM beta, suggesting that the liquidity premium is

at least partly systematic in nature. A simple calculation also suggests that the liquidity

returns on TIPS during normal times are comparable to the liquidity returns on o§-the-run

nominal Treasury bonds.

A possible interpretation of this result is that we partly identify a liquidity premium

or convenience yield on nominal Treasury bonds (Krishnamurthy and Vissing-Jorgensen,

2010), rather than a liquidity discount speciÖc to TIPS. Under this interpretation TIPS are

not undervalued securities; instead investors appear to be willing to pay a liquidity premium

on nominal Treasury bonds. The Treasury could take advantage of this premium by issuing

more nominal Treasury bonds, but it would still be issuing TIPS at their fair value. If

investors appropriately value TIPS, taking them o§ the market might have adverse welfare

consequences for investors in need of the real interest rate hedge and ináation hedge o§ered

by TIPS (Campbell and Viceira 2002).

Our results suggest several directions for future research. First, ináation expectations

are a major input into monetary policy. One could adjust breakeven ináation for the forms

of ináation risk premia and liquidity premia found in this paper to obtain a measure of longterm expected ináation. It would be informative to see whether this is a good predictor of

future ináation and other macroeconomic variables. Second, di§erent classes of investors have

di§erent degrees of exposure to time-varying liquidity, real interest rate risk and ináation risk.

It would be interesting to understand the implications for portfolio management and pension

investing and how these implications vary by investment horizon and the investorís share

of real and nominal liabilities. Third, our analysis of supply e§ects in the ináation-indexed

37
market suggests to further explore strategic behavior by the government in accommodating

shifts in the demand for nominal and real bonds.

38
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41
Table I

Summary Statistics

US data is monthly 1999.4-2009.12 and UK data is monthly 1985.4-2009.12. The US Asset-Swap

Spread (ASW) is monthly 2007.7-2009.4. The nominal government bond yield y

$

n;t, TIPS yield

y

T IP S

n;t ; breakeven ináation bn;t = y

$

n;t  y

T IP S

n;t ; O§-the-Run Spread, GNMA Spread and ASW are

expressed in annualized % units. The summary statistics for quarterly overlapping nominal excess

log returns xr$

n;t+1; TIPS excess log returns xrT IP S

n;t+1 and breakeven excess log returns xrb

n;t+1 are in

annualized %. Transaction volume refers to the log of TIPS transaction volume relative to the

transaction volume of all government coupon securities with at least 6 years to maturity.

Panel A: US 10 YR Panel B: UK 20 YR

Mean Std Min Max Mean Std Min Max

y

$

n;t 4:80 0:80 1:25 6:70 6:38 1:94 3:79 9:93

y

T IP S

n;t 2:55 0:85 1:25 4:29 2:79 1:14 0:56 4:57

bn;t 2:25 0:40 0:45 2:87 3:59 0:94 2:07 5:50

xr$

n;t+1 3:26 8:57 41:89 56:62 3:47 14:67 104:94 109:32

xrT IP S

n;t+1 4:16 7:67 65:02 56:26 1:66 9:25 67:90 64:47

xrb

n;t+1 0:91 7:22 39:52 74:24 1:81 11:97 103:21 84:79

O§-the-run Spr. 0:04 0:14 0:41 0:59 NA

GNMA Spr. 0:90 0:17 0:15 1:92 NA

Transaction Vol. 0:57 0:45 1:44 0:00 NA

ASW 0:64 0:37 0:44 1:30 NA

1
Table II

Breakeven onto Liquidity Proxies US

We regress the di§erence between nominal bond yields and TIPS (breakeven ináation) onto liquidity proxies. SP F denotes the 10 year

CPI- ináation forecast from the Survey of Professional Forecasters. Monthly (quarterly) regressions provide Newey-West standard errorswith three (four) lags in parentheses. The p-value of the F-test for no predictability is shown. * and ** denote signiÖcance at the 5%and 1% level, respectively.

(1) (2) (3) (4) (5) (6) (7) (8)

y$n;t  yT IP S n;t y$n;t  yT IP S n;t  SP F

O§-the-run Spr. 0:63 0:49 0:53 0:59 0:48 0:56 0:75 0:75

(0:23) (0:18) (0:18) (0:17) (0:14) (0:15) (0:19) (0:20)

GNMA Spr. 0:46 0:37 0:19 0:56 0:21 0:24 0:32

(0:08) (0:12) (0:11) (0:27) (0:16) (0:15) (0:15)

Transaction Vol. 0:16 0:28 0:30 0:32 0:23 0:22

(0:09) (0:08) (0:08) (0:12) (0:08) (0:09)

Asset-Swap-Spr. set 1 set 1 set 1 1:59 1:31 1:66 1:56

(0:20) (0:15) (0:21) (0:23)

(O§-the-run Spr.)2 488:7

(189:9)

(GNMA Spr.)2 171:9

(63:3)

const: 2::71 3:12 3:13 3:30 2:95 2:72 3:38 0:92

(0:04) (0:08) (0:08) (0:09) (0:11) (0:09) (0:12) (0:12)

p-value 0:00 0:00 0:00 0:00 0:00 0:00 0:00 0:00R2 0:44 0:56 0:64 0:67 0:70 0:47 0:71 0:70Sample 1999:3  2009:12 1999:3  2006:12 1999:Q1  2009:Q4

2
Table III

Breakeven Ináation onto Relative Supply

We regress the di§erence between nominal bond yields and ináation-indexed yields (breakeven ináation) onto measures of relative supplyWe control for liquidity measures as in Table II and a time trend. Supplyt denotes the amount of ináation-indexed bonds outstandingrelative to all nominal and ináation-indexed bonds outstanding. Supplyt denotes the issuance of ináation-indexed bonds relative to allnominal and ináation-indexed bonds issuance. "Supply t is obtained as the residual in a 12-lag monthly autoregression ofSupplyt  Supplyt12 (Supplyt) in the US (UK). Newey-West standard errors with three lags are provided in parentheses.

The p-value of the F-test for no predictability is shown. * and ** denote signiÖcance at the 5% and 1% level, respectively.

Panel A: US Panel B: UKSupplyt 0:01 0:01 0:27 0:29

(0:02) (0:04) (0:06) (0:05)

Supplyt 0:03 0:00 0:01 0:01

(0:02) (0:01) (0:01) (0:01)

"Supply t 0:04 0:00 0:00 0:00

(0:02) (0:01) (0:02) (0:01)

O§-the-run Spr. 0:57 0:60

(0:17) (0:15)

GNMA Spr. 0:19 0:20

(0:11) (0:09)

Transaction . 0:40 0:39

(0:15) (0:11)

Asset-Swap-Spr. 1:55 1:53

(0:22) (0:17)

month 105 0:11 0:42 3:43 1:96

(0:92) (0:32) (0:29) (0:21)

p-value 0:69 0:14 0:00 0:09 0:00 0:00 0:42 0:00 0:69 0:00

R2 0:01 0:04 0:68 0:04 0:70 0:26 0:00 0:66 0:00 0:53Sample 1999:3  2009:12 2000:2  2009:12 1985:1  2009:12 1986:1  2009:12

3
Table IV

Excess Bond Returns onto Relative Supply of Ináation-Indexed Bonds

We regress log excess bond returns of TIPS, nominal bonds and nominal bonds over TIPS onto measures of relative supply as deÖnedin Table III. We control for the nominal term spread, the TIPS term spread and the breakeven term spread as well as TIPS liquidity asestimated in Table II (4) and a time-trend. We use quarterly overlapping returns. Newey-West standard errors with three lags are providedin parentheses. The p-value of the F-test of no predictability is shown. * and ** denote signiÖcance at the 5% and 1% level, respectively.Panel A: US Panel B: UK

xr$n;t+1 xrT IP S n;t+1 xrbn;t+1 xrbn;t+1 xrbn;t+1 xr$n;t+1 xrT IP S n;t+1 xrbn;t+1 xrbn;t+1 xrbn;t+1

y$n;t  y$1;t 2:35 3:23

(1:52) (1:76)

yT IP S n;t  yT IP S 1;t 3:30 1:81 1:78 2:17 3:02 2:58

(1:23) (1:19) (1:25) (1:41) (1:95) (2:19)

bn;t  b1;t 5:67 6:44 7:00 7:39 8:97 9:00

(2:88) (2:75) (3:47) (2:27) (2:65) (2:85)

Ln;t 3:27 9:61 13:10 9:78 10:49

(6:49) (6:02) (3:78) (4:65) (4:27)

Supplyt 0:26 0:14 0:01 0:21 2:98 2:72 0:51 1:75

(0:70) (0:75) (0:62) (0:61) (2:30) (1:78) (2:25) (2:18)

Supplyt 0:17 0:10 0:17 0:04 0:14 0:12 0:00 0:05

(0:45) (0:36) (0:28) (0:25) (0:49) (0:34) (0:39) (0:41)

"Supply t 0:47 0:02

(0:37) (0:53)

month 104 1:27 0:73 1:83 1:69 1:33 2:05

(1:38) (1:41) (0:89) (1:13) (1:09) (0:69)

p-value 0:60 0:01 0:00 0:00 0:00 0:09 0:01 0:01 0:01 0:02R2 0:04 0:16 0:21 0:22 0:23 0:06 0:07 0:10 0:11 0:10Sample 1999:3  2009:12 1985:4  2009:12 1986:4  2009:12

4
Table V

Liquidity-Adjusted Return Predictability US

We regress liquidity-adjusted excess log bond returns of TIPS and of nominal

bonds in excess of TIPS onto the liquidity-adjusted TIPS term spread, the liquidity-adjusted

breakeven term spread and the liquidity premium on TIPS Ln;t. Ln;t is estimated as the

Ötted value less the constant from the regression in Table II (4). r

L

n;t+1 is the return on TIPS

due to liquidity. We use quarterly overlapping returns. Newey-West standard errors with three

lags appear in parentheses. The p-value of the F-test for no predictability is shown. * and **

denote signiÖcance at the 5% and 1% level, respectively.

(1) (2) (3) (4)

xrT IP SL

n;t+1 xrb+L

n;t+1 xrb+L

n;t+1 r

L

n;t+1



y

T IP S

n;t  Ln;t

 y

T IP S

1;t 3:53 1:66 0:03

(1:41) (1:12) (0:77)

(bn;t + Ln;t)  b1;t 2:94 3:42 2:62

(1:57) (1:53) (3:09)

Ln;t 6:05 0:73 2:00 19:81

(11:09) (6:23) (6:86) (6:18)

p-value 0:05 0:17 0:11 0:02

R2 0:06 0:04 0:07 0:26

Sample 1999:6  2009:12

5
Table VI

Moments of Realized and Fitted Bond Returns

We show summary statistics for realized excess log returns xrn;t and historical Ötted risk premia

Et (xrn;t+1). We show realized sample moments E^ (xrn;t) and ^ (xrn;t) for excess log returns

on nominal bonds, TIPS, breakeven, liquidity-adjusted breakeven, liquidity-adjusted TIPS and liquidity

returns. Betas are with respect to excess log stock returns including dividends. We obtain US

expected excess log returns for liquidity-adjusted breakeven, liquidity-adjusted TIPS and liquidity over

1999.6-2009.12 as Ötted values from Tables V (1), V (3) and V (4). Expected nominal, TIPS and

breakeven excess log returns breakeven were obtained in Páueger and Viceira (2011) from analogous

return-predictive regressions using no liquidity-adjustment over 1999.4-2009.12 (US) and 1985.4-

2009.12 (UK). Numbers shown are annualized (%). Newey-West standard errors for ^ are computed

with 3 lags. * and ** denote signiÖcance at the 5% and 1% level for ^, respectively.

Panel A: US E^ (xrn;t) ^ (xrn;t)  (Etxrn;t+1)



2

(Etxrn;t+1)

2(xr$

n;t)



2

(Etxrn;t+1)

2(xrT IP S

n;t )

Excess Log Return Nominal 3:26 0:40 1:56 3%

Excess Log Return TIPS 4:16 0:06 2:70 10% 12%

Excess Log Return BEI 0:91 0:47 3:24 14%

Liq.-Adj. Exc. Log Ret. BEI 0:74 0:07 1:38 3%

Liq.-Adj. Exc. Log Ret. TIPS 2:87 0:46 1:90 5% 6%

Log Return Liquidity 1:36 0:53 3:15 12%

Panel B: UK E^ (xrn;t) ^ (xrn;t)  (Etxrn;t+1)



2

(Etxrn;t+1)

2(xr$

n;t)



2

(Etxrn;t+1)

2(xrT IP S

n;t )

Excess Log Return Nominal 3:47 0:16 3:13 5%

Excess Log Return TIPS 1:66 0:15 1:84 2% 4%

Excess Log Return BEI 1:81 0:01 2:55 3%

6
-.5 0 . 5 1 1.5 2

Annualized (%)

-1.5 -1 -.5 0

Log Trading Vol

9 8 0 0 0 2 0 4 0 6 0 8 1 0

year

Log T rading Vol (Left) Off-the-Run (Right)

GNMA (Right) A sse t S wa p S p re a d (Ri g h t)

Figure1. US Liquidity Proxies 1999-2009. Relative TIPS Trading Volume from

New York Federal Reserve survey FR-2004; GNMA spread from Bloomberg;

10 Year On-the-Run O§-the-Run spread from the Federal Reserve and Bloomberg;

TIPS-Nominal Asset-Swap-Spread from Barclays Live.

1
5 0 100 150 200 250 9 8 0 0 0 2 0 4 0 6 0 8 1 0

year

Liquidity (bp) Li qui di ty (b p) Constrai ned

Liquidity (bp) Nonlinear

Figure 2. Estimated US Liquidity Premium. 10 Year TIPS Liquidity

Premium estimated in Table I (4). The constrained liquidity premium

constrainsthe coe¢ cient onto the Asset-Swap-Spread to  1 as in

Table I (3).The nonlinear liquidity premium with quadratic terms is

estimated in Table I (5).

2
0 1 2 3 4 9 8 0 0 0 2 0 4 0 6 0 8 1 0

year

BEI BEI Ad j usted

BEI Ad j usted No nl i nea r

Figure 3. Liquidity-Adjusted US 10 Year Breakeven Ináation.

We adjust breakeven ináation (the di§erence between nominal

and TIPS yields) for liquidity premia shown in Figure 2.

3
-1 0 1 2 3 4 9 8 0 0 0 2 0 4 0 6 0 8 1 0

year

yTIPS yT IPS Adj usted

yT IPS Adj usted Nonl i n ear

Figure 4. Liquidity-Adjusted US TIPS. We adjust US 10 Year

TIPS yields for liquidity premia shown in Figure 2.

4
0 5 1 0 1 5

DTIPS/D (%) (Right Axis)

. 5 1 1.5 2 2.5 3

US 10 YR Breakeven (%)

9 8 0 0 0 2 0 4 0 6 0 8 1 0

year

US 1 0 YR Breakeve n (%) DT IPS/D (%) (Ri gh t Axi s)

Figure 5A. US Relative Supply and 10 Year Breakeven Ináation.

Relative supply shows the total amount of TIPS relative

to the total amount of TIPS and nominal bonds outstanding.

5
8 1 0 1 2 1 4 1 6 1 8

DTIPS/D (%) (Right Axis)

2 3 4 5 6

UK 20 YR Breakeven (%)

8 5 9 0 9 5 0 0 0 5 1 0

year

UK 2 0 YR Breakeve n (%) DT IPS/D (%) (Ri gh t Axi s)

Figure 5B. UK Relative Supply and 20 Year Breakeven.

Relative supply shows the total amount of ináation-linked gilts relative

to nominal and ináation-linked gilts outstanding.

6
-10 0 1 0 2 0 3 0 9 9 0 0 0 1 0 2 0 3 0 4 0 5 0 6 0 7 0 8 0 9 1 0

year

Inflation RP (Annualized %) Real Rate RP (Annualized %)

Liquidity RP (Annualized %)

Figure 6A. Estimated US Risk Premia. The real rate risk

premium, ináation risk premium and liquidity return premium

are obtained as the Ötted expected excess returns for

liquidity-adjusted TIPS, liquidity-adjusted breakeven

and liquidity from Tables IV (1), IV (3) and IV (4).

7
-10 0 1 0 2 0 8 4 8 6 8 8 9 0 9 2 9 4 9 6 9 8 0 0 0 2 0 4 0 6 0 8 1 0

year

Nominal RP (Annualized %) TIPS RP (Annualized %)

Breakeve n RP (Annu al i zed %)

Figure 6B. Estimated UK Bond Risk Premia. The nominal

risk premium, TIPS risk premium and breakeven risk premium

were obtained for 1985.4-2009.12 in Páueger and Viceira (2011).

They were obtained as expected excess returns on nominal bonds,

ináation-indexed bonds and breakeven in return-predictive

regressions analogous to Table IV without liquidity-adjustments.

8
Appendix: An Empirical Decomposition of Risk and

Liquidity in Nominal and Ináation-Indexed

Government Bonds
A.I The Deáation Floor in TIPS Yields

In order to better understand the value of the deáation option in 10 year TIPS during the

crisis we compare our data series of 10 year TIPS yields with 10 year TIPS issued in 2007 and

2008. While in 2008 and 2009 these TIPS had very similar times to maturity they di§ered in

their reference CPIs. During the Önancial crisis the July 2018 was the most recently issued

10 year TIPS and was fully protected against deáation over its 10 year time to maturity.

The January 2017 TIPS on the other hand has a reference CPI that is 7% lower than the

reference CPI of the July 2018 TIPS. Hence, while the deáation option on the July 2018

TIPS was going to pay o§ if ináation was zero over the next 10 years the deáation option on

the January 2017 TIPS was only going to pay o§ if the price level declined substantially over

the next ten years and therefore it should reáect a much smaller deáation option or none at

all.

In Figure A.1 we show our series of 10 year TIPS yields together with the yields of the

10 year TIPS maturing in July 2018 with a reference CPI of 215.63997, the 10 year TIPS

maturing in January 2018 with a reference CPI of 209.49645 and the 10 year TIPS maturing

in January 2018 with a reference CPI of 201.66452 between 4/30/2008 and 4/9/2009. The

yields of individual TIPS issuances are from Bloomberg. We can see that during the Önancial

crisis the yield of the July 2018 TIPS was indeed lower than the yields of the January 2018

and January 2017 TIPS, possibly reáecting a deáation option. One could however also

imagine that, having been issued very recently, this issuance might have had a more active

investor base and therefore enjoyed better liquidity.

We can also see that the GSW zero-coupon ten year TIPS yields, which we use for our

1
main analysis, are much closer to the January 2018 and January 2017 TIPS yields during

the crisis than to the July 2018 TIPS yields, reáecting the parametric approach of GSW,

which smooths out idiosyncratic movements of yields of individual TIPS issuances.

A.II Nonlinear and Constrained SpeciÖcations for L

In this section we report additional robustness checks of our results for di§erent speciÖcations

of the liquidity premium L:

One might be concerned that our estimation of the liquidity premium L relies on extrapolation outside the range of historically observed data. The e§ect of our liquidity proxies

on the liquidity premium in TIPS might not be linear and this might lead to signiÖcant

errors in the extreme values of the estimated liquidity premium. The results in this section

are thought to complement the results in Table I in the paper, where we reported results

including quadratic terms.

We show the regressions including the squared terms for the o§-the-run spread and for

the GNMA spread in Table A.I. The asset-swap-spread and the transaction volume exhibit

too little variation over time and their squared terms do not enter signiÖcantly when added

together with the linear terms. The squared o§-the-run spread and the squared GNMA

spread enter signiÖcantly.

One might also think that interaction terms could matter. For example, one could imagine

that while transaction volume was low, TIPS were more a§ected by market-wide áight-toliquidity episodes and therefore the interaction term between transaction volume and the

2
o§-the-run spread should enter negatively. However, we see in our regressions in Table

A.I that the interaction terms between transaction volume, the o§-the-run spread and the

GNMA spread do not enter signiÖcantly.

Figures A.1 and A.2 show the liquidity-adjusted breakeven and liquidity-adjusted TIPS

yields with the liquidity premium estimated as in Table A.I (2) and in Table I (5). As an

additional robustness check we also show the corresponding quantities estimated as in Table

I (3), that is with the regression coe¢ cient on the asset swap spread constrained to equal

1:

Tables A.II and A.III report our return-predictive regressions from Table IV in the paper

using the nonlinear and the constrained estimate of L, respectively. Return-predictability

appears una§ected by the choice of speciÖcation.

Table A.IV reports means and standard deviations of ináation risk premia, real interestrate risk premia and liquidity premia. The ináation risk premia reported in Table A.IV are

slightly lower than in Table V in the paper, the liquidity premium is slightly lower and the

real interest rate risk premium is slightly higher but the general results are similar.

A.III Tradeable Excess Returns on TIPS

In our main analysis we consider the excess returns on TIPS over the return on the hypothetical real 3-month T-bill in order to capture real rate risk premia on long-term real

bonds over short-term real bonds. However, a real 3-month T-bill does not exist and one

might therefore interested in knowing whether the return-predictability regressions in Table

3
IV continue to hold when computing TIPS excess returns as nominal TIPS returns over the

nominal returns on a nominal 3-month T-bill and breakeven returns as the nominal TIPS

returns over the nominal return on a nominal bond. As one can see in Table A.V the results

are almost unchanged.

A.IV Newey-West Standard Errors with 12 Lags

Table A.VI reports the same regressions as Table IV in the paper but with Newey-West standard errors with 12 lags in brackets. If anything the additional lags increase the signiÖcance

of our coe¢ cients.

A.V Alternative Measures of Supply Shocks

In this section we construct an additional measures of supply shocks for the US and the UK.

One might think that governments aim for a long-term target value for the relative supply

of TIPS relative to nominals. In that case transitory deviations around a trend in relative

supply might be more appropriate measures of supply shocks.

Table A.VII reports Dickey-Fuller unit root tests for the relative supply Supplyt; relative issuance Supplyt and changes in relative issuance over the previous year Supplyt 

Supplyt12 for both the US and the UK. We include 12 lags in order to be able to capture

the annual issuance cycle.

In the US we cannot reject that the relative supply Supplyt follows a unit root process.

4
The same is true for relative issuance Supplyt

: This presumably indicates the regular annual

issuance schedule of nominal and ináation-indexed bonds in the US. The annual change in

the relative issuance Supplyt  Supplyt12; however, does not follow a unit root process.

This indicates that over our sample period in the US the issuance schedules are relatively

stable with shocks around a regular issuance schedule.

Table A.VIII reports the autoregressions for Supplyt and SupplytSupplyt12 with

twelve lags. While the coe¢ cients in the autoregression of relative issuance sum to greater

than one the coe¢ cients of the change in relative issuance sum to less than one.

Our analogous analysis in the UK reveals that we can reject a unit root process in relative

issuance Supplyt. That is, over our sample period in the UK relative supply appears to

áuctuate around a target level and with shocks to relative issuance. This agrees well with

the irregular issuance schedule of ináation-indexed gilts in the UK. We focus on data for

Supplyt and Supplyt starting in 1985. Data is available before 1985 but it is subject to

very large shocks during the Örst years of ináation-indexed gilt issuance and our yield data

only starts in 1985.

From this analysis we conclude that an appropriate measure of supply shocks in the US

should be the residual of the twelve month autoregression of Supplyt  Supplyt12. We

take our measure of relative supply shocks in the UK to be the residual in the twelve month

autoregression of relative issuance Supplyt: We Önd that these supply shocks do not enter

signiÖcantly either in Table II or in Table III. If instead we use the residuals from the other

autoregressions reported in Table A.VIII the results in Tables II and III remain unchanged.

5
A.VI Sub-Period Betas

In Table V in the paper we reported betas for the di§erent components of bond returns.

One might be interested whether these betas look similar over sub-periods that exclude the

Önancial crisis. In Table A.IX we therefore report betas for US returns over the shorter

period ending at the end of 2006. We also report the corresponding alphas for completeness.

We Önd that while the non liquidity-adjusted TIPS beta changes depending on whether the

Önancial crisis is included or not, the liquidity-adjusted betas are una§ected by the inclusion

of the Önancial crisis. These results conÖrm the importance of decomposing TIPS returns

into a liquidity component and a real interest rate return.

Over the full sample period liquidity returns appear to have a positive alpha that is

statistically signiÖcant from zero. However, one can also see that this alpha becomes statistically insigniÖcant over the more recent sub-period starting in 2002. This suggests that

the positive alpha was at least partly related to the novelty of TIPS and due to particular

market conditions during the Örst few years of TIPS issuance. Once TIPS were more solidly

established the intercept was still positive but no longer statistically signiÖcant.

For the UK we similarly report nominal, TIPS and breakeven betas for the periods 1985.3-

2006.12 and 1999.1-2009.12. The bond betas appear robust to the inclusion or exclusion

of the Önancial crisis period, consistent with our intuition that relative liquidity between

ináation-linked and nominal gilts should play less of a role in the UK. We look at the shorter

sub-period starting in 1999 as a way of checking whether UK government bond betas are

likely to have changed over time, as discussed in Campbell, Sunderam and Viceira (2011)

and Campbell, Shiller and Viceira (2009). Indeed we Önd that the government bond betas of

6
UK bonds look di§erent over the shorter time period, the breakeven beta being negative over

the most recent decade and the TIPS beta being smaller and insigniÖcant but still positive.

A.VII Factor Betas

Having seen that our liquidity variable exhibits systematic exposure to market returns, one

might be interested in whether it loads onto either the size factor (SMB) or the value factor

(HML) of Fama and French (1993) or onto the Pastor-Stambaugh liquidity factor of Pastor

and Stambaugh (2003). We obtain monthly data for the SMB, HML and Pastor-Stambaugh

factors from CRSP. For the Pastor Stambaugh factor we use the innovations series. We

construct quarterly overlapping returns by summing over the past three months.

We Örst regress our bond excess returns onto market excess returns and the PastorStambaugh factor, which is a measure of changes in stock market liquidity. In Table A.X

we Önd that both nominal and ináation-indexed excess returns are negatively related to the

Pastor-Stambaugh factor. This holds for the full sample period and when we exclude the

Önancial crisis. Liquidity-adjusted TIPS returns also appear negatively related to the PastorStambaugh factor. One possible explanation is that when stock market liquidity decreases

investors rush into government bonds, driving up the prices of both nominal and ináationindexed Treasuries. Liquidity returns do not appear related to the Pastor-Stambaugh factor.

In Table A.XI we add the value and size factors to the regressions. Liquidity returns

do not appear related to either value or size. Even though both of these factors are enter

insigniÖcantly, the Pastor-Stambaugh factor now becomes signiÖcant but not if we exclude

the Önancial crisis period. Given the results in Table A.X the relationship between liquidity

7
returns and the Pastor-Stambaugh factor does not appear robust.

A.VIII Correlations of Returns and Spreads

One might be interested in how correlated are the returns and right-hand side variables in

Table V. These correlations are reported in Table A.XI. As one can see, the correlations of our

right-hand side variables in Table V are only weakly correlated, suggesting that collinearity

of regressors should not need to be a concern.

8
Table A.I

Liquidity Regression with Interaction Terms

We add interaction terms to the regressions in Table II. Newey-West standard errors with three lags in parentheses.

The p-value of the F-test for no predictability is shown. * and ** denote signiÖcance at the 5% and 1% level.

y

$

n;ty

T IP S

n;t (1) (2) (3) (4) (5)

O§-the-run Spr. 0:49 0:48 0:20 0:54 0:46

(0:14) (0:14) (0:33) (0:24) (0:14)

GNMA Spr. 0:18 0:56 0:52 0:560:58

(0:11) (0:27) (0:27) (0:28) (0:28)

Transaction Vol. 0:28 0:30 0:29 0:30 0:131

(0:09) (0:08) (0:08) (0:08) (0:26)

Asset-Swap-Spr. 1:49 1:31 1:32 1:30 1:39

(0:18) (0:15) (0:15) (0:15) (0:15)

(O§-the-run Spr.)

2 560:9

 488:7 438:7 464:2 499:0

(171:7) (189:9) (160:2) (205:3) (192:7)

(GNMA Spr.)

2 171:9

 162:1 174:8 157:8

(63:3) (63:93) (64:17) (56:19)

O§-the-run Spr.GNMA Spr. 110:7

(108:6)

O§-the-run Spr.Transaction Vol. 45:45

(124:0)

GNMA Spr.Transaction Vol. 73:34

(100:9)

const: 3:27 2:94 2:95 2:94 2:94

(0:08) (0:11) (0:11) (0:11) (0:11)

p  value 0:00 0:00 0:00 0:00 0:00

R2 0:69 0:70 0:70 0:70 0:71

Sample 1999:3  2009:12

1
Table A.II

Liquidity-Adjusted Predictability US

Nonlinear Liquidity Estimate

We run liquidity-adjusted return-predictive regressions as in Table V. Variables are

deÖned as in Table V. We adjust for the liquidity premium with nonlinear terms

as estimated in Table II(5). Newey-West standard errors with three lags appear in

parentheses. The p-value of the F-test for no predictability is shown. * and **

denote signiÖcance at the 5% and 1% level, respectively.

xrT IP SL

n;t+1 xrb+L

n;t+1 xrb+L

n;t+1 r

L

n;t+1



y

T IP S

n;t  Ln;t

 y

T IP S

1;t 4:07 2:12 0:51

(1:48) (0:99) (0:80)

(bn;t + Ln;t)  b1;t 3:07 3:63 2:41

(1:58) (1:51) (3:40)

Ln;t 8:47 2:26 3:57 21:74

(8:11) (3:77) (4:35) (4:62)

p  value 0:01 0:10 0:03 0:00

R2 0:08 0:07 0:12 0:28

Sample 1999:6  2009:12

Table A.III

Liquidity-Adjusted Predictability US

Constrained Liquidity Estimate

We run liquidity-adjusted return-predictive regressions as in Table V. Variables are

deÖned as in Table V. We adjust for the liquidity premium estimated in Table II (3)

with coe¢ cient on the asset-swap spread constrained to -1. Newey-West standard

errors with three lags appear in parentheses. The p-value of the F-test for no

predictability is shown. * and ** denote signiÖcance at the 5% and 1% level, respectively.

xrT IP SL

n;t+1 xrb+Ln;t+1 xrb+Ln;t+1 r

L

n;t+1



y

T IP S

n;t  Ln;t

 y

T IP S

1;t 3:22 1:70 0:40

(1:49) (1:15) (0:79)

(bn;t + Ln;t)  b1;t 3:92 4:30 2:38

(1:80) (1:74) (2:34)

Ln;t 3:56 1:63 0:53 16:87

(9:48) (5:42) (6:39) (4:90)

p  value 0:10 0:09 0:07 0:01

R2 0:05 0:07 0:10 0:23

Sample 1999:6  2009:12

2
Table A.IV

Moments of Realized and Fitted Bond Returns

Nonlinear and Constrained Estimates

We show summary statistics for expected and realized excess returns as in Table VI for di§erent

speciÖcations of the liquidity premium. We obtain expected excess returns 1999.6-

2009.12 as Ötted values from Tables A.II (1), (3) and (4) and Tables A.III (1), (3) and (4),

respectively. Numbers shown are annualized (%).

Panel A: Nonlinear Liquidity Estimate

E^ (xrn;t) ^ (xrn;t)  (Etxrn;t+1)



2

(Etxrn;t+1)

2(xr$

n;t)



2

(Etxrn;t+1)

2(xrT IP S

n;t )

Liq.-Adj. Exc. Log Ret. BEI 0:23 0:06 3:37 4%

Liq.-Adj. Exc. Log Ret. TIPS 3:38 0:46 4:69 7% 9%

Log Return Liquidity 0:86 0:52 6:99 19%

Panel B: Constrained Liquidity Estimate

E^ (xrn;t) ^ (xrn;t)  (Etxrn;t+1)



2

(Etxrn;t+1)

2(xr$

n;t)



2

(Etxrn;t+1)

2(xrT IP S

n;t )

Liq.-Adj. Exc. Log Ret. BEI 0:57 0:04 3:25 4%

Liq.-Adj. Exc. Log Ret. TIPS 3:04 0:36 3:29 4% 4%

Log Return Liquidity 1:20 0:42 4:74 9%

3
Table A.V

Return Predictability Liquidity US

Tradeable Excess Returns

We run liquidity-adjusted return-predictive regressions as in Table V. We consider

liquidity-adjusted log nominal bond returns on TIPS over nominal returns on the

nominal T-bill rate r

T IP SL

n;t  r

T bill

t

as a tradeable version of liquidity-adjusted log

excess returns of TIPS. We also consider liquidity-adjusted log nominal bond returns

on TIPS over log nominal bond returns on nominal bonds r

T IP SL

n;t  r

nom

n;t as a

tradeable version of liquidity-adjusted breakeven returns. Newey-West standard errors

with three lags appear in parentheses. The p-value of the F-test for no predictability is

shown. * and ** denote signiÖcance at the 5% and 1% level, respectively.

r

T IP SL

n;t  r

T bill

t

r

T IP SL

n;t  r

nom

n;t r

T IP SL

n;t  r

nom

n;t



y

T IP S

n;t  Ln;t

 y

T IP S

1;t 3:56 1:71

(1:44) (1:21)

(bn;t + Ln;t)  b1;t 3:06 3:56

(1:76) (1:71)

Ln;t 7:67 2:26 3:57

(10:77) (6:27) (7:00)

p  value 0:05 0:17 0:14

R2 0:06 0:05 0:07

Sample 1999:6  2009:12

Table A.VI

Return Predictability Liquidity US

Newey-West Standard Errors with 12 Lags

We run liquidity-adjusted return-predictive regressions as in Table V.

Newey-West standard errors with twelve lags appear in parentheses.

The p-value of the F-test for no predictability is shown. * and ** denote

signiÖcance at the 5% and 1% level, respectively.

xrT IP SL

n;t+1 xrb+L

n;t+1 xrb+L

n;t+1 r

L

n;t+1



y

T IP S

n;t  Ln;t

 y

T IP S

1;t 3:53 1:66 0:03

(1:43) (0:92) (0:9)

(bn;t + Ln;t)  b1;t 2:94 3:42 2:62

(1:41) (1:31) (3:52)

Ln;t 6:05 0:73 2:00 19:81

(6:80) (3:98) (4:46) (5:27)

p  value 0:04 0:07 0:05 0:00

R2 0:06 0:04 0:07 0:26

Sample 1999:6  2009:12

4
Table A.VII

Dickey-Fuller Unit Root Tests

We report Dickey-Fuller test for the null-hypothesis of a unit root with twelve

lags. Supplyt denotes the amount of ináation-indexed bonds outstanding

relative to all nominal and ináation-indexed bonds outstanding. Supplyt

denotes the relative issuance of ináation-indexed bonds relative to all nominal

and ináation-indexed bonds issuance. Supplyt  Supplyt12 is the

change in relative issuance over the past 12 months. * and ** denote signiÖcance

at the 5% and 1% level respectively.

Panel A: US

Supplyt Supplyt Supplyt  Supplyt12

2:03 1:72 3:05

Sample 1999:1  2009:12 1999:2  2009:12 2000:2  2009:12

Panel B: UK

Supplyt Supplyt Supplyt  Supplyt12

1:67 4:33 1:88

Sample 1986:1  2009:12 1986:1  2009:12 1987:1  2009:12

Table A.VIII

Supply Autoregression

We regress relative issuance Supplyt; as described in Table III onto its own twelve lags. We also regress

the 12-month change in relative issuance onto its own twelve lags. We report the sum of the twelve

autoregressive coe¢ cients.

Panel A: US Panel B: UK

Supplyt Supplyt

Supplyt Supplyt12 Supplyt Supplyt12

Sum of Coe§. 1:10 0:92 0:47 1:05

const: 0:00 0:00 0:00 0:00

(0:00) (0:00) (0:00) (0:00)

R2 0:69 0:30 0:15 0:28

Sample 1999:2  2009:12 2000:2  2009:12 1986:1  2009:12 1987:1  2009:12

5
Table A.IX

Sub-Period Betas US

We regress liquidity-adjusted and non liquidity-adjusted excess log government bond returns

onto excess log stock returns. All variables are described in Table VI. Newey-West standard errors are computed

with 3 lags. * and ** denote signiÖcance at the 5% and 1% level for ^ and ^, respectively.

Panel A: US 1999.4-2009.12 1999.4-2006.12 2002.1-2009.12

b b b b b b

Excess Log Return Nominal 0:40 2:52 0:51 2:47 0:38 3:46

Excess Log Return TIPS 0:06 4:28 0:34 3:74 0:11 4:50

Excess Log Return BEI 0:47 1:76 0:17 1:28 0:49 1:03

1999.6-2009.12 1999.6-2006.12 2002.1-2009.12

b b b b b b

Liq.-Adj. Exc. Log Ret. BEI 0:07 0:87 0:02 0:72 0:09 0:97

Liq.-Adj. Exc. Log Ret. TIPS 0:46 1:96 0:51 2:17 0:47 2:49

Log Return Liquidity 0:53 2:40 0:17 1:62 0:58 2:00

Panel B: UK 1985.3-2009.12 1985.3-2006.12 1985.3-1998.12 1999.1-2009.12

b b b b b b b b

Excess Log Return Nominal 0:16 2:98 0:19 3:25 0:31 3:96 0:06 0:41

Excess Log Return TIPS 0:15 1:21 0:14 1:13 0:17 0:09 0:13 2:40

Excess Log Return BEI 0:01 1:78 0:05 2:12 0:13 3:87 0:19 1:98

6
Table A.X

Loadings onto the Market and the Pastor-Stambaugh Factor

US liquidity-adjusted and non liquidity-adjusted excess log government bond returns onto excess log

stock returns xr

equity

t

and the innovations in the Pastor-Stambaugh liquidity factor. Annualized (%).

Newey-West standard errors with 3 lags in brackets. * and ** denote signiÖcance at the 5% and 1% level.

Panel A: 1999.6-2009.12 xr$

n;t+1 xrT IP S

n;t+1 xrbn;t+1 xrT IP SL

n;t+1 xrb+Ln;t+1 r

L

n;t+1

xr

equity

t+1 0:28 0:19 0:47 0:40 0:12 0:59

(0:13) (0:24) (0:19) (0:16) (0:12) (0:16)

Pastor-Stambaugh 0:09 0:10 0:01 0:05 0:04 0:05

(0:03) (0:03) (0:04) (0:03) (0:02) (0:02)

const. 2:42 3:91 1:49 1:73 0:70 2:18

(1:69) (1:52) (1:35) (1:52) (1:24) (1:12)

Panel B: 1999.6-2006.12 xr$

n;t+1 xrT IP S

n;t+1 xrb

n;t+1 xrT IP SL

n;t+1 xrb+Ln;t+1 r

L

n;t+1

xr

equity

t+1 0:39 0:27 0:13 0:45 0:06 0:18

(0:22) (0:14) (0:14) (0:13) (0:13) (0:04)

Pastor-Stambaugh 0:11 0:08 0:03 0:06 0:05 0:01

(0:03) (0:02) (0:02) (0:02) (0:02) (0:01)

const. 2:81 3:74 0:93 2:12 0:69 1:61

(2:01) (1:34) (1:36) (1:28) (1:43) (0:50)

7
Table A.XI

Four Factor Regressions

We add the Fama-French factors SMB and HML to the regression in TableA.X. Annualized (%).

Newey-West standard errors with 3 lags in brackets. * and ** denote signiÖcance at the 5% and 1% level.

Panel A: 1999.6-2009.12 xr$

n;t+1 xrT IP S

n;t+1 xrb

n;t+1 xrT IP SL

n;t+1 xrb+Ln;t+1 r

L

n;t+1

xr

equity

t+1 0:25 0:21 0:47 0:37 0:12 0:58

(0:13) (0:25) (0:20) (0:17) (0:11) (0:16)

HML 0:05 0:00 0:05 0:01 0:06 0:00

(0:05) (0:08) (0:06) (0:06) (0:04) (0:03)

SMB 0:13 0:10 0:03 0:10 0:03 0:01

(0:06) (0:06) (0:05) (0:06) (0:05) (0:03)

Pastor-Stambaugh 0:09 0:10 0:00 0:05 0:04 0:05

(0:03) (0:03) (0:03) (0:03) (0:02) (0:02)

const. 3:08 4:67 1:59 2:56 0:52 2:11

(1:73) (1:72) (1:42) (1:58) (1:11) (1:17)

Panel B: 1999.6-2006.12 xr$

n;t+1 xrT IP S

n;t+1 xrb

n;t+1 xrT IP SL

n;t+1 xrb+Ln;t+1 r

L

n;t+1

xr

equity

t+1 0:33 0:23 0:10 0:42 0:09 0:18

(0:23) (0:15) (0:14) (0:14) (0:13) (0:04)

HML 0:06 0:09 0:03 0:09 0:03 0:01

(0:06) (0:04) (0:04) (0:04) (0:04) (0:02)

SMB 0:10 0:01 0:10 0:00 0:10 0:01

(0:06) (0:04) (0:05) (0:04) (0:05) (0:02)

Pastor-Stambaugh 0:11 0:08 0:03 0:06 0:04 0:01

(0:03) (0:03) (0:02) (0:03) (0:02) (0:01)

const. 3:22 2:86 0:36 1:35 1:87 1:51

(2:28) (1:58) (1:20) (1:43) (1:28) (0:54)

8
Table A.XII

Sample Correlations of Returns and Spreads

Monthly data of quarterly overlapping returns and ináation 1999.6-2009.12.

Annualized (%). All data are described in Table V.

Correlations Returns

xrT IP SL

n;t+1 xrb+Ln;t+1 r

L

n;t+1

xrT IP SL

n;t+1 1 0:24 0:44

xrb+L

n;t+1 : 1 0:21

r

L

n;t+1 : : 1

Correlations Spreads



y

T IP S

n;t  Ln;t

 y

T IP S

1;t (bn;t + Ln;t)  b1;t Ln;t



y

T IP S

n;t  Ln;t

 y

T IP S

1;t 1 0:25 0:18

(bn;t + Ln;t)  b1;t : 1 0:27

Ln;t : : 1

9
Figure A.1. Recently Issued and Less Recently Issued 10 Year TIPS. GSW 10 Year

TIPS yields from Gurkaynak, Sack and Wright (2010); 10 Year TIPS maturing in 07/2018

(reference CPI 215.7), 01/2018 (reference CPI 209.5) and 01/2017 (reference CPI 201.7)

from Bloomberg.

10
0 1 2 3 4

98 00 02 04 06 08 10

year

BEI BEI Adjusted

BEI Adjusted Constrained BEI Adjusted Nonlinear

Figure A.2. Constrained and Nonlinear Liquidity Adjustments to Breakeven.

We adjust breakeven ináation (the di§erence between nominal and TIPS yields) for

liquidity premia shown in Figure 2. Relative to Figure 3 we add breakeven ináation

adjusted for the constrained liquidity estimate.

11
-1 0 1 2 3 4

98 00 02 04 06 08 10

year

yTIPS yTIPS Adjusted

yTIPS Adjusted Constrained yTIPS Adjusted Nonlinear

Figure A.3. TIPS Yields Adjusted for Constrained and Nonlinear Liquidity SpeciÖcations.

We adjust US 10 Year TIPS yields for liquidity premia shown in Figure 2. Relative to Figure 4

we add TIPS yields adjusted for the constrained liquidity estimate.

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