Market based inflation forecast

Carlo A Favero

February 2016

Contents

1 Model Based, Survey Based and Asset Prices based Forecast of Inflation 2

1.1 Asset Pricing with Time-Varying Expected Returns . . . . . . . . . . . . . . 3

2 Market-based Instruments to Forecast Inflation 5

2.1 Inflation indexed bonds . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 5

2.1.1 Further technical details . . . . . . . . . . . . . . . . . . . . . . . . . 6

2.2 Inflation Swaps . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 7

2.2.1 Further technical details . . . . . . . . . . . . . . . . . . . . . . . . . 7

3 Bond Returns: Yields-to-Maturity, Duration and Holding Period Returns 7

3.1 Zero-Coupon Bonds . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 8

3.2 A simple model of the term structure . . . . . . . . . . . . . . . . . . . . . . 9

3.3 Forward Rates . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 9

3.3.1 Instantaneous forward rates . . . . . . . . . . . . . . . . . . . . . . . 10

4 Financial Factor Models of the Term Structure: 11

5 A general state-space representation 12

6 Assessing the importance of the risk premium 12

6.1 Single-equation evidence on the ET . . . . . . . . . . . . . . . . . . . . . . . 12

6.2 The VAR based evidence . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 16

7 No-Arbitrage Affine Factor Models 18

7.1 Risk Premium . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 20

7.2 Extensions . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 21

8 Appendix: Coupon Bonds 22

8.1 Zero-Coupon Equivalent . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 22

8.2 Log-linearization . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . . 23

1. Model Based, Survey Based and Asset Prices based Forecast of Inflation

The object of our interest is forecasting inflation. As an illustrative case consider the case of

US Consumer Price Index for All Urban Consumers: All Items, seasonally adjusted observed

at monthly frequency over the sample 1999-2015.

Graphical analysis of the data shows clearly that inflation at different frequency (monthly,

quarterly and annual) has diffferent time-series properties in terms of persistence and volatil-

ity.

As well discussed by Ang, Bekaert and Wei(2006), economists use four main methods to

forecast inflation. The first method is atheoretical, using time series models of the ARIMA

variety. The second method builds on the economic model of the Phillips curve, leading to

forecasting regressions that use real activity measures. Third, inflation can be forecasted

using survey based measures that use information from agents (consumers or profession-

als) directly to forecast inflation. Finally, using information embedded in asset prices, in

particular the term structure of interest rates, can be used to forecast inflation.

Each method has its own specific problems.

Univariate ARIMA methods and multivariate VAR based approaches derived by

reduced form of economic models are backward-looking. In addition, in the case of theoretical

model a choice must be made on the frequencuy of the data that is more consistent with

the generic ”period” in the economic model, time-series methods have to make choices on

the seasonal adjustment and on the level of aggregation at which the data are used in the

forecasting model. Suppose the object of interest is annual inflation. Is it better to forecast

directly the aggregate or to forecast inflation at higher frequencies and then aggregate them

to obtain an annual forecast ?

Survey based forecast reflect preferences of the forecasters. If the loss function is not

symmetric, then inflation based forecast are different from expected inflation. Clatworthy et

al. (2005) illustrate very well this point by showing that biased forecast are not necessarily

irrational. Consider for example forecasters that have a LINEX loss function:

= − − 1

2

where is a constant and is the forecast error at time t. The parameter determines

the degree of asymmetry. A LINEX function with = 07 is plotted in Figure 1:

2

If 0, the Linex function is exponential to the right of the origin, and linear to the left.

In this case, pessimistic forecasts (positive forecast errors) are more costly than optimistic

forecasts (negative forecast errors). Conversely when 0 optimistic forecasts (negative

forecast errors) are more costly than pessimistic forecasts (positive forecast errors) . In this

case, the loss is approximately exponential in x if x 0, and approximately linear in x if x

0.

Under the LINEX loss function the optimal forecast is

And therefore the problem of extracting expected inflation from surveys becomes more

complicate as identification bites.

Market based measures of expected inflation are forward looking, and are based on the

action of agents who put their money at the stake. However, also in the case of market

bases measures there is a relevant identification problem to which we shall devote a section

as market based measures are the main focus of our attention.

1.1. Asset Pricing with Time-Varying Expected Returns

Different approaches have been used in finance to model time-varying expected returns, they

are all understood within the context of a basic model that stems from the assumption of

the absence of ”arbitrage opportunities” (i.e. by the impossibility of making profits without

3

taking risk). Consider a situation in which in each period k state of nature can occur and

each state has a probability () in the absence of arbitrage opportunities the price of an

asset i at time t can be written as follows:

=

X=1

+1()+1 ()+1 ()

where +1 () is the discounting weight attributed to future pay-offs, which (as the

probability ) is independent from the asset i, +1 () are the payoffs of the assets ( in

case of stocks we have +1 = +1 ++1 in case of zero coupon bonds, +1 = +1),

and therefore returns on assets are defined as 1 + +1 =+1

For the safe asset, whose

payoffs do not depend on the state of nature, we have:

= +1

X=1

+1()+1 ()

1 ++1 =1

P=1

+1()+1 ()

In general, we can write:

= (+1+1)

1 ++1 =1

(+1)

consider now a risky asset :

(+1 (1 ++1)) = 1

(+1+1) = 1− (+1) (1 ++1)

(1 ++1) = − (+1+1)

(+1)+ (1 ++1)

Turning now to excess returns we can write:

(+1 −+1) = − (1 ++1) (+1+1)

Assets whose returns are low when the stochastic discount factor is high (i.e. when

agents values payoffs more) require an higher risk premium, i.e. an higher excess return

on the risk-free rate. Turning to predictability at different horizon, if you consider the

case in which t is defined by taking two points in time very close to each other the safe

4

interest rate will be approximately zero and will not vary too much across states. The

constant expected return model (with expected returns equal to zero) is compatile with the

no-arbitrage approach at high-frequency. However, consider now the case of low frequency,

when t is defined by taking two very distant points in time; in this case safe interest rate

will be diffferent from zero and will vary sizeably across different states. The constant

expected return model is not a good approximation at long-horizons. Predictability is not a

symptom of market malfunction but rather the consequence of a fair compensation for risk

taking, then it should reflect attitudes toward risk and variation in market risk over time.

Different theories on the relationship between risk and asset prices should then be assessed

on the basis of their ability of explaining the predictability that emerges from the data.

Also, different theories or return predictability can be interpreted as different thoeries

of the determination of On the one hand we theories of based on rational investor

behaviour, on the other hand we have alternative approaches based on psycological models

of investor behaviour. Our main interest is to show how the predictability of returns can be

used for optimal portfolio allocation purposes, rather than on discriminating between the

possible sources of predictability.

2. Market-based Instruments to Forecast Inflation

2.1. Inflation indexed bonds

The most common form of government bonds are nominal bonds that pay fixed coupons

and principal. However, at the end of the last millenium governments have started issuing

significant amounts of inflation-indexed bonds. Inflation-indexed bonds, which in the U.S.

are known as Treasury Inflation Protected Securities (TIPS), are bonds whose coupons and

principal adjust automatically with the evolution of a consumer price index. They aim to

pay investors a fixed inflation-adjusted coupon and principal, in other words they are real

bonds and their yields are typically considered the best proxy for the term structure of real

interest rates in the economy.

Investors holding either inflation-indexed or nominal government bonds are exposed to

the risk of changing real interest rates.In addition to real interest rate risk, nominal govern-

ment bonds expose investors to inflation risk while real bonds are not. When future inflation

is uncertain, the coupons and principal of nominal bonds can suffer from the eroding effects

of inflationary surprises. If inflation is negatively correlated with economic conditions, as in

times of stagflation, the real payoffs of nominal bonds will tend to decline when economic

conditions worsen. Risk averse investors will therefore demand a positive inflation risk pre-

mium for hold-ing nominal bonds. But if inflation is positively correlated with economic

5

conditions, nominal bonds will have hedging value to risk-averse investors. Finally, both the

nominal and real bond are theoretically affected by a premium for liquidity risk. Liquidity

risk, is, the risk of having to sell (or buy) a bond in a thin market and, thus, at an unfair

price and with higher transaction costs. In practice small issuers with low volumes of bonds

outstanding and thus small markets must compensate investors with a liquidity premium

and in the case of US Treasuries we can assume that the liquidity premium will only affect

the TIPS.

At time the yields to maturity of nominal and real bonds maturing at can be wriiten

as follows:

= + +

+

= +

+

the difference in the yield to maturity, usually referred to as the breakeven inflation rate

can be written as:

= + +

from which it is clear that expected inflation is equivalent to the breakeven infaltion rate

only when the effect of the sum of inflation risk and liquidity risk premium can be neglected.

2.1.1. Further technical details

Further details complicate the pricing of real bonds. Inflation figures are published with a

lag and therefore the principal value of inflation indexed bonds is adjusted with a lag (three-

month in the US and the UK). Moreover the mecahanism of indexation could be different

for coupons and pricncipal and it might have different provisions for positive and negative

inflation. Finally the tax treatment of the bond might also differ. In the UK principal

adjustments of inflation-linked gilts are not taxed. This gives inflation-linked gilts a tax

advantage over nominal gilts, a larger share of whose cash flows come in the form of taxable

nominal coupon payments. In the US, on the other hand, inflation-adjustments of principal

are considered ordinary income for tax purposes. As a result the tax obligations associated

with holding a TIPS increase when inflation is high so that on an after-tax basis. In practice

U.S. TIPS are not fully indexed to inflation.

6

2.2. Inflation Swaps

An inflation swap is a bilateral contractual agreement. It requires one party (the ‘inflation

payer’) to make periodic floating-rate payments linked to inflation, in exchange for prede-

termined fixed-rate payments from a second party (the ‘inflation receiver’). Inflation swap

contracts are arranged OTC so the pay-off structure can be matched to the needs of the

counterparty. The most common structure is the zero-coupon inflation swap. The inflation

receiver pays on a contract maturing in T years pays the Fixed leg = (1 + fixed rate)T

x Nominal value to receive from the the inflationpayer the variable Inflation leg = (Final

price index/Starting price index) x Nominal value. Therefore, in the case of indexation to

contemporaneous inflation the return on the fixed leg of a swap contract which is normally

referred to as inflation compensation can be written as :

= +

2.2.1. Further technical details

In practice, inflation swap contracts have indexation lags. This means a contract is referenced

to inflation for a period that begins periods before the date on which the contract is priced

and ends before the contract matures.

= −− +

As the object of interest is inflation from time t onward, the following decomposition

becomes relevant:

=

− − +

−−

− − +

which, in absence of publication lags on inflation data allows to derive the relevant

inflation compensation from the fixed swap rate.

3. Bond Returns: Yields-to-Maturity, Duration and Holding Period Returns

Analyzing bonds is crucial to understand how information on expected inflation can be

extracted bythe term structure. We distinguish between two type of bonds: those paying a

coupon each given period and those that do noy pay a coupon but just reimburse the entire

capital upon maturity (zero-coupon bonds). We shall work mainly with zero-coupon bonds

but all our results can be extended to coupon bonds.

Cash-flows from different type of bonds:

7

+ 1 + 2 + 3

general +1 +2 +3

coupon bond 1 +

1-period zero 1 0 0 0

2-period zero 0 1 0 0

(T—t) -period zero 0 0 0 1

3.1. Zero-Coupon Bonds

Define the relationship between price and yield to maturity of a zero-coupon bond as follows:

=1

(1 + )− (1)

where is the price at time of a bond maturing at time , and is yield to maturity.

Taking logs of the left and the right-hand sides of the expression for and defining the

continuously compounded yield, as log(1 + ), we have the following relationship:

= − ( − ) (2)

which clearly illustrates that the elasticity of the yield to maturity to the price of a zero-

coupon bond is the maturity of the security. In other words the duration of the bond equals

maturity as no coupons are paid.

Table 2.:Price and YTM of zero-coupon bonds

Maturity 1 2 3 5 7 10 20

09524 09070 08638 07835 07106 06139 03769

00500 00500 00500 00500 00500 00500 00500

−00487 −00976 −01464 −02439 −03416 −04879 −09757 00488 00488 00488 00488 00488 00488 00488

The one-period uncertain holding-period return on a bond maturing at time , +1 is

then defined as follows:

+1 ≡ +1 − = − ( − − 1) +1 + ( − ) (3)

= − ( − − 1) (+1 − )

= ( − ) − ( − − 1) +1 (4)

which means that yields and returns differ by the a scaled measure of the change between

the yield at time + 1 +1 and the yield at time , . Think of a situation in which

the one-year YTM stands at 4.1 per cent while the 30-year YTM stands at 7 per cent. If

the YTM of the thirty year bonds goes up to 7.1 per cent in the following period, then the

period returns from the two bonds is the same.

8

3.2. A simple model of the term structure

Apply the no arbitrage condition to a one-period bond (the safe asset) and a T-period bond:

¡+1 − 1+1

¢=

¡+1 − +1

¢= +1

¡+1

¢= +1 + +1

Solving forward the difference equation = +1 − +1 we have :

=1

( − )

−1X=0

¡+++1

¢=

1

( − )

−1X=0

¡+++1 + +++1

¢Which makes clear that YTM on long bond depends on future expected returns on one-

period bond and the average risk premium over the maturity of the long bond. The simplest

model of the term structure, the expectations theory, posits that the average risk premium

is constant.

3.3. Forward Rates

Forward rates are returns on an investment at time , made in the future at time 0 with

maturity at time . The return on this strategy is equivalent to the return on a strategy

that buys at time t zero coupon with maturity and sells at time t the same amount of

bonds with maturity 0

The price of the investment strategy is (− ( − ) + (0 − ) 0)and using the usual

formula that links prices to returns we have :

0 =( − ) − (0 − ) 0

− 0(5)

Applying the general formula to specific maturities we have :

+1+2 = 2+2 − +1 (6)

+2+3 = 3+3 − 2+2 (7)

+3+4 = 4+4 − 3+3 (8)

9

+−1+ = + − (− 1)+−1 (9)

Using all these equations we have:

+ =1

(+1 + +1+2 + +2+3 + + +−1+) (10)

+ =1

−1X=0

¡+++1 + +++1

¢therefore each spot rate can be interpreted as an average of future one period forward.

Putting together this evidence with the model for the term structure we have:

+++1 =

¡+++1 + +++1

¢Think of using forward rates to assess the impact of monetary policy. Let us analyze a

potential movement of spot and forward rates around a shift in the central bank target rate.

Before CB intervention1-year spot and forward rates

maturity i=1 i=2 i=3 i=4 i=5

+ 0.05 0.05 0.05 0.05 0.05

+++1 0.05 0.05 0.05 0.05

After CB intervention :1-year spot and forward rates

maturita i=1 i=2 i=3 i=4 i=5

+ 0.06 0.06 0.05 0.045 0.04

+++1 0.06 0.03 0.03 0.02

Please remember that the interpretion of future forward as expected rates requires some

assumption on the risk premium.

3.3.1. Instantaneous forward rates

Define the instanteous froward as the forwad rate on the contract with infinitesimal maturity:

0 = lim→0

0 (11)

given the sequence of forward rates you can define forward rate at any settlement date as

follows :

0 =

R =0

( − 0)

10

As a consequence the relationship between spot and forward rate is written as:

=

R =

( − )

and therefore

= + ( − )

(12)

so instantaneous forward rates and spot rates coincide at the very short and very long-

end of the term structure, forward rates are above spot rates when the yield curve slopes

positevely and forward rates are below spot rates when the yield curve slopes negatively.

4. Financial Factor Models of the Term Structure:

The empirical financial literature has put the concept just illustrated at work to build model

of the term structure that can be used to interpolate nontraded maturity and to forecast

yield at each maturity.

Gurkanyak et al. estimate the following interpolant at each point in time, by non-linear

least squares, on the cross-section of yields:

+ = +

1− exp³−

1

´1

+ 1

⎛⎝1− exp³−

1

´1

− expµ−

1

¶⎞⎠+ (13)

+2

⎛⎝1− exp³−

2

´2

− expµ−

2

¶⎞⎠which is an extension originally proposed by Svensson(1994) on the original parameterization

adopted by Nelson and Siegel (1987) that sets 2 = 0

Forward rates are easily derived as

= + exp

µ−

1

¶+ 1

1exp

µ−

1

¶+ 2

2exp

µ−

2

¶(14)

When maturity goes to zero forward ans spot rates coincide at + and when maturity

goes to infinite forward and spot coincide at . Terms in 1 and 2

describes two humps

starting at zero at different starting points and ending at zero. As discussed in Diebold

and Li (2002) the above interpolant is very flexible and capable of accommodating several

stylized facts on the term structure and its dynamics. In particular, 1

2 which

are estimated as parameters in a cross-section of yields, can be interpreted as latent factors.

has a loading that does not decay to zero in the limit, while the loading on all the other

parameters do so, therefore this parameter can be interpreted as the long-term factor, the

11

level of the term-structure. The loading on is a function that starts at 1 and decays

monotonically towards zero; it may be viewed a short-term factor, the slope of the term

structure. In fact, = + is the limit when k goes to zero of the spot and the

forward interpolant. We naturally interpret as the risk-free rate. Obviously , the

slope of the yield curve, is nothing else than the minus the spread in Campbell-Shiller.

are medium term factor, in the sense that their loading start at zero, increase and then decay

to zero (at different speed). Such factors capture the curvature of the yield curve. In fact,

Diebold and Li show for example that 1 tracks very well the difference between the sum

of the shortest and the longest yield and twice the yield at a mid range (2-year maturity).

The repeated estimation of loadings using a cross-section of eleven US yields at different

maturities allows to construct a time-series for four factors.

5. A general state-space representation

To generalize the NS approach we can put the dynamics of the term structure in a state-space

framework. Yields with different maturities are collected in a vector = [+1 +2 +]0.

Equation (15) is the measurement equation, in which different yields + are assumed to

be determined by a set of state variables, collected in the vector :

+ =−1( +0

) + + ∼ (0 2) (15)

= + Φ−1 + ∼ (0Ω) (16)

In the case of original NS we have

0 =

∙− −

µ1− −

¶−µ1− −

− −

¶¸and = 0

Note that factors can be first extracted in the cross-sectional dimension and then their

time-series behaviour can be studied in a VAR or the model can be estimated simultaneously,

using an unobservable component method such as the Kalman filter.

6. Assessing the importance of the risk premium

We consider two approaches to assessing the importance of the risk premium: single equation

based-evidence and VAR based evidence.

6.1. Single-equation evidence on the ET

The traditional evidence is mainly based on a single-equation, limited information, approach.

12

To assess the importance of RP, three different implications of the Expectations theory

can be brought to the data:

− ( − − 1) (+1 − ) = +1 + (17)

+++1 =

¡+++1 + +++1

¢(18)

+ − +1 =

−1X=1

µ1− 1

¶∆+++1 +

1

−1X=1

+++1(19)

(a) Estimate the following model :

+1 − = 0 + 11

− − 1 ( − +1) + +1

to test 0 = 0 1 = 1

(b) Estimate the following model :

∗ = 0 + 1 +

= + − +1

∗ =

−1X=1

µ1− 1

¶∆+++1

to test 0 = 0 1 = 1

(c) Estimate the following model :

(+++1 − +1) = 0 + 1 (+++1 − +1) + ++1

to test 0 = 0 1 = 1

(d) Estimate the following model :

− ( − − 1) (+1 − )− +1 = 0 + 1 (+++1 − +1) + ++1

to test 1 = 0

The empirical evidence shows that:

• i) high yields spreads fare poorly in predicting increases in long rates(see Campbell,1995)

• ii) the change in yields does not move one-to-one with the forward spot spread (seeFama and Bliss,1986)

• iii) period excess returns on long-term bond are predictable using the information in

the forward-spot spread (see Cochrane,1999,Cochrane-Piazzesi 2005)

13

14

15

6.2. The VAR based evidence

The VAR based evidence is due to Campbell-Shiller(1987) and considers the following

version of the ET

= ∗ =−−1X=1

[∆+ | ] (20)

(20) shows that a necessary condition for the ET to hold puts constraints on the long-run

dynamics of the spread. In fact, the spread should be stationary being a weighted sum of

stationary variables. Obviously, stationarity of the spread implies that, if yields are non-

stationary, they should be cointegrated with a cointegrating vector (1,-1). However, the

necessary and sufficient conditions for the validity of the ET impose restrictions both on the

long-run and the short run dynamics.

Assuming1 that and are cointegrated with a cointegrating vector (1,-1), CS con-

struct a bivariate stationary VAR in the first difference of the short-term rate and the spread

:∆ = ()∆−1 + ()−1 + 1

= ()∆−1 + ()−1 + 2(21)

Stack the VAR as:⎡⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎣

∆

∆−+1

−+1

⎤⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎦=

⎡⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎣

1 1

1 0 0 0

0 0 0 0

0 1 0 0 0

1 1

0 0 1 0

0 0 0 0

0 0 0 1 0

⎤⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎦

⎡⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎣

∆−1

∆−−1

−

⎤⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎦+

⎡⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎢⎣

1

0

2

0

⎤⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎥⎦(22)

This can be written more succinctly as:

= −1 + (23)

The ET null puts a set of restrictions which can be written as :

0 =−1X=1

00 (24)

1In fact, the evidence for the restricted cointegrating vector which constitutes a necessary condition for

the ET to hold is not found to be particularly strong in the original CS work.

16

Figure 1:

where 0 and 0 are selector vectors for and ∆ correspondingly ( i.e. row vectors with 2p

elements, all of which are zero except for the p+1st element of 0 and the first element of 0which are unity). Since the above expression has to hold for general , and, for large T, the

sum converges under the null of the validity of the ET, it must be the case that:

0 = 0( − )−1 (25)

which implies:

0( − ) = 0 (26)

and we have the following constraints on the individual coefficients of VAR(21):

= −∀ 1 = −1 + 1 = −∀ 6= 1 (27)

The above restrictions are testable with a Wald test. By doing so using US data between

the fifties and the eighties Campbell and Shiller (1987) rejected the null of the ET. However,

when CS construct a theoretical spread ∗ by imposing the (rejected) ET restrictions on

the VAR they find that, despite the statistical rejection of the ET, ∗ and are strongly

correlated.

17

7. No-Arbitrage Affine Factor Models

The standard response in finance to the empirical rejection of the Expectations Theory

has been modelling the term structure based on the assumption that there are no riskless

arbitrage opportunities among bonds of various maturities. The standard model is based on

three components: a transition equation for the state vector relevant for pricing bonds, made

traditionally of latent factors, an equation which defines the process for the risk-free one-

period rate and a relation which associates the risk premium with shocks to the state vector,

defined as a linear function of the state of the economy. In such structure, the price of a j-

period nominal bond is a linear function of the factors. Unobservable factors and coefficients

in the bond pricing functions are jointly estimated by maximum likelihood methods (see, for

example, Chen and Scott(1993)). This type of models usually provides a very good within

sample fit of different yields but do not perform well in forecasting. Duffee(2002) shows

that the forecasts produce by no-arbitrage models with latent factors do not outperform the

random walk model.

First building block is the dynamics of the factors determining risk premium:

= + Φ−1 + Σ

interpret this as a companion form representation.

Second is a specification for the one-period rate which is assumed to be a linear function

of the factors:

= 0 + 01

The third is a pricing kernel. The assumption of no-arbitrage guarantees the existence of

a risk-neutral measure Q such that the price of any asset that does not pay any dividends

at time t+1 satisfies the following relation:

= (exp (−)+1)

the Radon-Nikodym derivative (which converts the risk neutral measure to the data-

generating measure) is denoted by +1 So for any random variable +1 we have

(+1) =

¡+1+1

¢

The assumption of no-arbitrage allows us to price any nominal bond in the economy.

Assume that +1 follows the log-normal process:

+1 = exp

µ−120 − 0+1

¶18

where are the time-varying market prices of risk associated with the sources of uncer-

tainty Parameterize as an affine process:

= 0 + 1

define the pricing kernel +1 as

+1 = exp (−) +1substituting from the processes for the short-rate and +1 we have:

+1 = exp

µ−0 − 01 − 1

20 − 0+1

¶Now the total one-period gross return of any nominal asset satisfies:

(+1+1) = 1

If represents the price of an n-period zero coupon bond, then we can use this equation

to compute recursively bond prices as:

+1 =

¡+1

+1

¢Now guess that the log of bond prices are linear functions of the state variable we have:

= exp ( +0)

This guess is easily verified for the one-period bond, in which case we have:

1 = (+1) = exp (−)= exp (0 + 01)

But it also applies to n-period bonds, in which case we have:

19

+1 =

¡+1

+1

¢= exp

µ−0 − 01 − 1

20 − 0+1 + +0

+1

¶= exp

µ−0 − 01 − 1

20 +

¶ [exp (−0+1 +0

+1)]

= exp

µ−0 − 01 − 1

20 +

¶ [exp (−0+1 +0

(+ Φ + Σ+1))]

= exp

µ−0 + (0

Φ− 01) − 120 + +0

¶ [exp (−0+1 +0

Σ+1)]

= exp

µ−0 + (0

Φ− 01) + +0 (− Σ0) +

1

20ΣΣ

0 −0Σ1

¶here the last step uses log-normality and the fact that 0 = 0 (+1)By matching

coefficients we now have:

+1 = −0 + +0 (− Σ0) +

1

20ΣΣ

0

0+1 = 0

(Φ− Σ1)− 01

To sum up, we can characterize a traditional Affine TS model as follows:

+ =−1( +0

) + + ∼ (0 2) (28)

= + Φ−1 + ∼ (0Ω) (29)

+1 =1

(+1)

∙P=0

(Φ0 − 01Ω)

¸1

+1 = 1 − 1(+1)

P=0

(), where () = 0 (−Ω0) +

120Ω.

7.1. Risk Premium

Given the knowledge of the model parameters the risk premium can be derived naturally:

= + − 1

−1X=0

+

+ = 0 + 01+

= 0 + 01¡+ Φ ( − )

¢20

where

= + Φ−1 + Σ

= ( −Φ)−1

and, in absence of measurement error, we have:

+ =−1( +0

)

= +0

= −

X 0

− 01

à − 1

−1X=0

Φ

!+

+

Ã0 − 01

1

−1X=0

Φ

!

but 0+1 = 0

(Φ−Σ1)− 01so when Σ = 0 or 1 = 0 the term multiplying in the

last expression vanishes and the risk premium becomes constant.

7.2. Extensions

Recently the no-arbitrage approach has been extended to include some observable macroeco-

nomic factors in the state vector and to explicit allow for a Taylor-rule type of specification

for the risk-free one period rate. Ang and Piazzesi(2002) and Ang, Piazzesi and Wei(2003)

show that the forecasting performance of a VAR improves when no-arbitrage restrictions are

imposed and that augmenting non-observable factors models with observable macroeconomic

factors clearly improves the forecasting performance. Hordahl et al.(2003) and Rudebusch

and Wu(2003) use a small scale macro model to interpret and parameterize the state vector;

forecasting performance is improved and models have also some success in accounting for

the empirical failure of the Expectations Theory.

This extension of the small information set using macroeconomic variable can be further

expanded by moving to large-information set. In this case rather than including in the state

vector some specific macroeconomic variables, common factors can be extracted from a large

panel of macroeconomic variables using static principal components, as suggested by Stock

and Watson (2002).

21

8. Appendix: Coupon Bonds

The relationship between price and yield to maturity of a constant coupon () bond is given

by:

=

¡1 +

¢ + ¡1 +

¢2 + +1 +

(1 + )−

To apply the framework we ahve considered so far to Coupon Bonds therea are two alterna-

tives.

1) Find the zero-coupon equivalent

2) Consider log-linearization

8.1. Zero-Coupon Equivalent

Define the discount function at time of a zero-coupon paying one unit at + :

+ = exp (−+) (30)

=1

(1 + +) = + (31)

Consider now a coupon-bond:

+ =

X=1

+ ++ (32)

Coupon bonds are nothing else than a bundle of zero coupons. For coupon bonds youa have

available yield to maturity computed, given the price, as :

+ =

X=1

exp¡−+¢++

The tow formula can be used to build a zero coupon equivalent curve, given that you have

available at least one zero coupon for the shortest maturity. the methodology is recursive

and it is called ”bootstrapping” :

+ =

⎛⎜⎜⎝ 1 + +

1−P−1=1

+

(1++)

⎞⎟⎟⎠1

− 1

The following table illustrates an application

YTM and zero coupon equivalent

YTM +1 +2 +3 +4 +5 +6 +7 +8 +9

4.69 4.64 4.72 4.82 4.92 5.01 5.10 5.17 5.23

zero-coupon equivalent +1 +2 +3 +4 +5 +6 +7 +8 +9

4.69 4.64 4.72 4.83 4.94 5.04 5.14 5.22 5.29

22

Given the one-year zero coupon the zero coupon equivalent forthe two year is generated

as follows :

+2

1 + +2

+1 + +2³1 + +2

´2 = +1

1 + +2

+1 + +2

(1 + +2)2

from which we have:

+2 =

⎛⎝ 1 + +2

1− +21++1

⎞⎠12

− 1

Having obtained +2the equation :

+3

1 + +3

+ +3³

1 + +3

´2 + 1 + +3³1 + +3

´3 = +3

1 + +2

+1 + +3

(1 + +2)2+

1 + +3

(1 + +3)3

can be solved for +3 :

+3 =

⎛⎜⎜⎝ 1 + +3

1−P2

=1

+3

(1++)

⎞⎟⎟⎠13

− 1

by iteration the full term structure is then derived

8.2. Log-linearization

When the bond is selling at par, the yield to maturity is equal to the coupon rate. To

measure the length of time that a bondholder has invested money for we need to introduce

the concept of duration:

=

(1+ )

+ 2

(1+ )

2 + + ( − ) 1+

(1+ )−

=

−P=1

(1+ )

+(−)

(1+ )−

23

Note that when a bond is floating at par we have:

=

−X=1

¡1 +

¢ + ( − )

(1 + )−

=

³( − ) 1

1+

− ( − )− 1´

1

(1+ )

−+1 +1

1+ ³

1− 11+

´2 +( − )

(1 + )−

=1− ¡1 +

¢−(−)1− ¡1 +

¢−1

because when || 1X

=0

=(− − 1)+1 +

(1− )2

Duration can be used to find approximate linear relationships between log-coupon yields

and holding period returns. Applying the log-linearization of one-period returns to a coupon

bond we have:

− = −+1 + + (+1 − )

+1 = + +1 + (1− ) −

When the bond is selling at par, = (1 + )−1=¡1 +

¢−1. Solving this expression

forward to maturity delivers:

=

−−1X=0

¡ + (1− ) − +1+

¢

The log yield to maturity satisfies an expression with the same structure:

=

−−1X=0

¡ + (1− ) −

¢=1− −−1

1−

¡ + (1− ) −

¢=

¡ + (1− ) −

¢

By substituting this expression back in the equation for linearized returns we have the

expression

+1 =

−

¡

− 1¢+1

that illustrates the link between continuously compounded returns and duration.

A simple model of the term structure of coupon bonds Consider the relation be-

tween the return on a riskless one period short-term bill, and a long term bond bearing

a coupon the one-period return on the long-term bond is a non-linear function of

24

the log yield to maturity Shiller (1979) proposes the linearization illustrated in the

previous section which takes duration as constant and considers the following approximation

in the neighborhood = +1 = = :

' − ( − 1) +1

=1− −−1

1− =

1

1−

=n1 +

£1− 1(1 + )−−1

¤−1o−1lim

−→∞ = = 1(1 + )

solving this expression forward :

=

−−1X=0

(1− )+ + −−1

In this case, by equating one-period risk-adjusted returns, we have:

∙ − +1

1− | ¸= + (33)

From the above expression, by recursive substitution, under the terminal condition that

at maturity the price equals the principal,we obtain:

= ∗ +[Φ | ] = 1−

1− −

−−1X=0

[+ | ] +[Φ | ] (34)

where the constant Φ is the term premium over the whole life of the bond:

Φ =1−

1− −

−−1X=0

+

For long-bonds, when − is very large, we have :

= ∗ +[Φ | ] = (1− )

−−1X=0

[+ | ] +[Φ | ]

Subtracting the risk-free rate from both sides of this equation we have:

= − =

−1X=1

[∆+ | ] +[Φ | ]

= ∗ +[Φ | ]

25