AR(1) time series with autoregressive gamma variance for road topography modeling

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1 Working Papers in Statistics No 5:5 Department of Statistics School of Economics and Management Lund University AR() time series with autoregressive gamma variance for road topography modeling PÄR JOHANNESSON, SP TECHNICAL RESEARCH INSTITUTE OF SWEDEN KRZYSZTOF PODGÓRSKI, LUND UNIVERSITY IGOR RYCHLIK, CHALMERS UNIVERSITY OF TECHNOLOGY NIMA SHARIATI, LUND UNIVERSITY

4 .5 Probability density of slopes..5 PDF Slope [%] Figure : Models of slopes pdf; the Gaussian pdf N(, 4.8) (dashed line), the generalized Laplace pdf defined in () with parameters m =, ν =.8, σ = 4.8 (solid line) and the normalized histogram. representation of a generalized Laplace variable Y = σ RX + m, () where X the standard normal (Gaussian) distribution and is independent of gamma distributed factor R that has mean one and variance ν defined by the density function f(r) = Γ (/ν) ν /ν r ν e r/ν, () where Γ( ) is the gamma function. The non-gaussianity (shape) parameter ν can be estimated using ν = κ 3, (3) 3 where κ is the kurtosis, which for the Gaussian case is κ = 3. This parameter has a simple interpretation as the excess of kurtosis (κ 3) measured in terms of the Gaussian kurtosis. Note also that if the non- Gaussianity parameter ν tends to zero, the random variance R converges (in distribution) to a constant factor equal one and consequently, Y becomes N(m, σ ) in the limit. The mean and variance of a Laplace variable are also given respectively by E [Y ] = m and V [Y ] = σ. (4) We refer to [] for a comprehensive treatment of the generalized Laplace distributions. Although the Laplace distribution often fits the empirical records quite well, the constant variance over data segments has an additional difficulty in the need to impose a model for the lengths of the constant variance segments. The simplest model assuming equally spaced segments and independent variances usually cannot be adopted since it is observed, not surprisingly, that variances of adjacent parts of the record are typically correlated. To circumvent this problem effectively, we use the above ideas by building a simple model that does not require ad hoc splitting of the data into segments. Namely, it is reasonable to believe that the quality of road as measured by variance varies slowly and hence the variances are likely dependent between themselves. It would then be appropriate to model them by means of an autocorrelated model. Hence, a combination of an autoregressive gamma model for variances with a classical Gaussian autoregressive models results in a flexible non-gaussian time series defined by only five parameters, mean m, variance σ, shape ν and two autoregressive parameters, ρ x and ρ r. Empirical example - slopes of a road in a mine The presented Laplace time series model will be illustrated using measured slopes of a 33 km long segment of rough road surface in a mine. The slopes are measured in % and sampled each meters. 3

5 The data are presented in Figure. In the figure, sections with almost constant variance are marked by the vertical lines. The signal has mean zero, variance 4.8 and kurtosis 6.3. In Figure the normalized histogram of the measured slopes is compared with the probability density functions (pdf) of the normal model N(, 4.8) and the generalized Laplace pdf defined in (5) with parameters m =, σ = 4.8 and ν =.8. We observe that the Laplace distribution adequately represents the empirical distribution of the data. As it is shown in the next sections, variances of slopes can be fitted fairly well by the autoregressive gamma process. 3 Autoregressive gamma variance AR() model The key property of the autoregressive gamma variance model is that it introduces the dependence to the random gamma variance scaling in () with a single parameter that control the degree of the dependence. The two features of the model need to be emphasized: firstly, it preserves the generalized Laplace distribution of the data; secondly, it replaces the concept of segments with constant variances by introducing the dependence between varying variances. In this way, the problematic issue of identifying intervals of constant variance is replaced by straightforward estimation of a single parameter that controls the lag of dependence. It is worth to mention that the model has also extension to account for asymmetry in data distribution. Here is how the symmetric model for the slopes Y k is formally introduced, in which we follow, Y k = σ R k X k + m (5) where X k s follow the Gaussian AR(), while R k s are gamma autoregressive random variables independent of X k s. The model was first considered in [8], where it was used in a model of parallel tracks. These two components of the model are described below in further details. Figure 3 illustrates the capacity of the model to account for varying variance. In the top graph, we can see a simulated record with segments of clearly different variabilities, resembling records in Figure. In the bottom graph, we see the corresponding random variance itself. In our application, the gamma factors R k s reflect the variability of the road hilliness while the Gaussian process X k models the local variability of the slopes. It should be emphasized that our restriction to the AR() model for X k is because of simplicity and one can also consider, for example, higher order of autoregressive Gaussian processes. The model will be used to describe the variability of the slopes of a road in a mine shown in Figure. Due to the intended simplicity, the proposed stationary time series model does not attempt to capture all aspects of the measured signals. Our focus is on modeling the long-term distribution of some properties relevant for fuel consumption, e.g. length, height and average steepness of hills. Autoregressive gamma model for R k In our application, the gamma factors R k s model the variability of the variance of slopes in Y k. The quality of a surface varies slowly and hence the factors R k s are likely dependent between consecutive parts of the road. As reported in [8], the stationary gamma process R k, with correlation coefficient ρ r and shape ν can be written as R k = ρ r K(R k ) R k + ( ρ r ) ɛ k, (6) where ρ r (, ) is an autoregressive coefficient, ɛ k s are mutually independent gamma distributed variables random innovations with mean one and variance equal to the parameter ν, independent also from K k = K(R k ) and R k. Here the random factor K(r) is given by K(r) = N(r) E i, (7) m(r) i= 4

6 3 Simulated model Y Autoregressive gamma variance 8 6 R Figure 3: Top: Simulated values of Y. Bottom: Corresponding simulated values of autoregressive gamma variance. where E i s are independent random variables, E = while, for i >, E i s are exponentially distributed and independent of the Poisson random variable N(r) having mean m(r) equal to m(r) = ρ r r ρ r ν. (8) It has been shown that this model has exponentially decaying autocorrelation ρ k r, k =,,..., see [6], where the gamma autoregressive model has been introduced, or [], where a historical overview and further properties of this model are presented. Gaussian AR() time series for X k While an arbitrary stationary Gaussian time series X k leads to a generalized Laplace distribution for Y k, we restrict ourself to the autoregressive X k of order one with the autocorrelation function ρ X (k) = ρ k x, ρ x (, ), normalized so that variance equals to one. We have the following classical recursive formula X k = ρ x X k + ρ x ε k, (9) where ε k s are independent zero mean Gaussian variables with variance one (Gaussian white noise). The parameter ρ x completely determines the joint distribution of X k s. We note that the vector process (X k, R k ) has the Markov property, while Y k, defined in (5) as a function of (X k, R k ) is not a Markov process, anymore. 4 Autocorrelation functions One approach to estimation of the parameters that govern correlation structures is to consider sample correlation estimates and matching them with the theoretical ones. For this one has to derive the formulas for autocorrelations of some functions defined on the process. For our model two autocorrelation functions are of interest: ρ y (k) = corr ( Yk, Y ) and ρy (k) = corr (Y k, Y ), k =,,,.... In the 5

7 case of process Y k centered at zero, Yk can be viewed as a local estimate of the variance, and thus ρ y (k) should roughly represent the dependence of the random variance R k. Thus, the autocovariances of the squared records should serve well for estimation of the parameter ρ r. On the other hand, the covariance of the Gaussian part of the model, i.e. the process X k should be more evident in the covariance of ρ y (k). These claims are formalized in the following result, where we provide explicit formulas for autocorrelations, while their derivation is elaborated in the appendix. Proposition. For the model Y k given in (5) with R k and X k following (6) and (9), respectively, we have (i) the autocorrelation of non-negative process Y k ρ y (k) = ρk x (ii) the autocorrelation of process Y k is given by where and ρ r /(k) is the autocorrelation of R / k. is given by + (ρ k x + )ν ρ k r + 3ν. () ρ y (k) = (a ν + ( a ν )ρ r /(k)) ρ k x, () a ν = ν Γ ( ν + ) Γ ( ν ) () The autocorrelation ρ r /(k) of R k depends on ρ r and also on ν, so it can be formally written as a certain function h(ρ r, ν, k), ρ r (, ), ν >, k =,,.... A general explicit form of h(ρ r, ν, k) is rather difficult to obtain. Here we present only the case k = : ) Γ ( ν + (( ) ρ r ) ν + F ν h(ρ r, ν, ) = +, ν +, (ρ r) ν Γ( ν )Γ( ν + ) Γ ( ν + ) (3) = a ( ) ν ( ρ r ) ν + F a ν ν +, ν +, (ρ r), ν where F a,b,c (z) is a hypergeometric function defined as F (a, b; c; z) = n= (a) n (b) n (c) n where (q) n is the (rising) Pochhammer symbol. We note that for most practical purposes, one can use a simpler approximate relation. Namely, Figure 4 (left) shows the behavior of h(ρ r, ν, ) = ρ r /() in terms of different ν and ρ r. Since the deviation of ρ r /() from ρ r is very small (with the maximal error of approximation reaching.464), one can simply use ρ r instead of ρ r /(), with the approximation error illustrated in Figure 4 (Right). From (), in the case of k =, we get z n n!, ρ y () = a ν ( ρ r ) ν + F ν +, ν +, ν (ρ r) ρ x, where a ν is given in (). However, we saw in Figure 4 that there is a quite range of parameters which this relation can be approximated by ρ y () = (a ν + ( a ν )ρ r )ρ x. (4) Finally, we note that while both X k and R k have AR()-type correlations, the autocorrelation function ρ y (k) of Y k is not of an exponential form ρ k x except in the case when ρ r =. In this latter case ρ y (k) = a ν ρ k x, if k >. (5) In this special case, there is a jump in autocorrelation at k = and this property can be used for detection if the simplified model is applicable to the data. 6

8 .5 h( r,,)= r / r r / r r Figure 4: Left: Graph of h(ρ r, ν, ) = ρ r /() as a function of ν and ρ r. Right: Graph of difference of ρ r /() and ρ r. 5 Estimation of model parameters The model (5) has all together five parameters, the center m, the scale σ, the shape parameter ν and two one-step correlations, ρ r and ρ x. Estimation of the center m and the scale σ can be done in the standard way, i.e. by taking the sample mean and sample standard deviation. From now on, we assume that our data are standardized so we can assume that m = and σ = and our goal is to estimate the remaining three parameters: the shape parameter ν and the two correlations, ρ x and ρ r. Parameter ν can be obtained from matching sample excess kurtosis with ν via (3), which reduces the problem to estimation of the two one-step correlations, ρ r and ρ x. 5. Autocorrelation method Fitting ρ x and ρ r can be obtained through fitting the covariances of Y and Y. The method is based on the equations () and (7). The formula for the one lag autocorrelation of Y can be used to establish a relation between ρ x and ρ r. Namely, for a sample estimate ˆρ y of one step correlation of the data, we can set the equation ˆρ y = (( â)h(ρ r ) + â) ρ x, where â = aˆν and h(ρ) = h(ρ, ˆν, ) is given by (3). To make the relation even simpler, one can use approximated relation (4) to obtain the explicit relation Let us now introduce a function G k (ρ) = ν ρ x = ˆρ y â + ( â)ρ r. (6) ( + (3ν + )ρy (k) ρ k + ). (7) Based on (), the function G k should be equal to ρ k r when it is evaluated at ρ = ρ x. Consequently, we have the possibility of using (7) to match ρ r with the empirical evidence, by replacing ρ y by the standard sample estimate of the autocorrelation of the squared values of the signal. Then, using the right hand side of (7), one can use sample estimates ˆρ y (k) of ρ y (k) to obtain functions Ĝk(ρ x ), k =,..., K of the yet undetermined ρ x. The value of the maximal lag K should be chosen not too large so that at this lag there is still some meaningful dependence in the data. Since these functions should be approximately equal to ρ k r, we set them equal to one when Ĝk(ρ x ) exceeds one and to zero when it is smaller than zero. Further, if we use (6), this function becomes a function of ρ r, denoted now by H k (ρ r ): ( ) ˆρ y H k (ρ r ) = Ĝk. (8) â + ( â)ρ r 7

9 An estimate of ρ r can be found by solving in ρ: ρ k = H k (ρ). The solutions, say ˆρ rk, can be combined through a weighted geometric average with linearly decreasing weights ˆρ r = (ˆρ K r ˆρ K r... ˆρ rk ) K(K+). Finally, ˆρ r can be plugged into (6) to get the corresponding estimate ˆρ x of ρ x. We note that the above fitting method is suitable when the Gaussian component is characterized by one parameter, as it is the case in our AR() model. For more general Gaussian components, one can alternatively use the one-step correlation formula for Y k to relate ρ r with ρ x () and then use () in the appendix to fit ρ x (k), for more general models of ρ x (k). 5. Zero-crossing method The Gaussian component X is characterized by a single parameter ρ x. Since in our process the factors Rk are strictly positive, the observed zero crossings coincide with the zero crossings of X; the property that can be used to estimate ρ x. In other words, we can utilize an important characteristic of the time series of slopes the frequency of encountered hills f (in the sampling frequency unit) is defined as follows. Let N n be the number of encountered hills, i.e. the number of times that Y k up-crosses level zero for k =,..., n. Then the (dimensionless) frequency is defined as N n f = lim n n = lim n #{i =,..., n : Y i <, Y i+ > }, n where #A denotes number of elements in a set A. For the slopes Y k, defined in (5), the frequency of hills is equal to the frequency of zero upcrossings in the Gaussian time-series X k. Now, for a zero mean stationary and ergodic Gaussian time series, the frequency of zero upcrossings is given by f = P(X <, X > ) = 4 π arcsin ρ x, (9) which is the special case α =, β = of a general relation for standardized jointly Gaussian variables X and Y with correlation ρ: ( ) [ E X +α Y β] Γ α+β + α+β ( = ρ ) π/ α+β+ cos φ sin β φ dφ, π ( ρ sin(φ)) α+β + that can be found in []. Here x + = max(, x) while x = max(, x). One can use (9) by taking ˆρ x matching the empirically observed frequency giving ˆf = #{i =,..., n : Y i <, Y i+ > }, n ˆρ x = sin(π( ˆf.5)). () Once the estimation of ρ x is concluded, one can substitute the estimate of ρ x in () and estimate ρ r by fitting it to the empirical correlation ˆρ y (k). Namely, the resulting ˆρ rk = ] /k [Ĝk (ˆρ x ), k =,..., K, are used to match ρ r. There are many ways to do so, since we have a sequence of values and only one parameter ρ r to match them. Since the accuracy of these estimators depends and generally decreases 8

10 Table : Estimation of ν, ρ x and ρ r for two trajectory lengths: 5 roman, and italics. The numbers represent Monte Carlo approximation of the mean values of the estimates (in the parentheses are the standard deviation of the estimates) based on independent samples of size and lag K = 3 Estimators Parameters Autocorrelation mehtod Zero-crossing method ν ρ x ρ r ˆν(ˆσˆν ) ˆρ x ˆρ r ˆρ x ˆρ r (.).6 (.4).57 (.9).6 (.4).58 (.). (.).6 (.).7 (.6).6 (.).69 (.5).87 (.5).6 (.3).83 (.6).59 (.3).87 (.).96 (.).6 (.).89 (.8).6 (.).89 (.9).87 (.).8 (.).63 (.).78 (.3).66 (.). (.).79 (.).74 (.6).8 (.).7 (.6).98 (.4).8 (.5).8 (.6).8 (.).85 (.).5 (.6).8 (.).88 (.7).8 (.).9 (.6).76 (.43).58 (.5).68 (.3).59 (.3).68 (.4).4 (.6).6 (.).7 (.8).6 (.).7 (.7).84 (.49).6 (.4).8 (.).6 (.3).86 (.).9 (.8).6 (.).88 (.7).59 (.).9 (.6).9 (.44).78 (.6).69 (.5).75 (.).74 (.4).95 (.8).79 (.3).69 (.9).78 (.).7 (.7).58 (.53).8 (.5).79 (.3).78 (.4).84 (.).95 (.39).79 (.).9 (.6).79 (.).9 (.5) with k, the fitting procedure should account for this effect. Here, we take a simple approach again using the weighted geometric average with linearly decreasing weights to combine the solutions ˆρ r = (ˆρ K r ˆρ r K ) K(K+)... ˆρ rk. () It is worth to note that the method described in the previous subsection did not use the observed frequency of hills which were used here to estimate ρ x. Consequently, by combining the previous approach with the zero-crossing method one can, in principle, fit models for which autocorrelation of X involves two parameters. Such more general models are not considered in this work. 5.3 Comparisons of the estimation methods For verification of the proposed estimation procedures, we simulate data points from the model using different combinations of parameters ν, ρ x and ρ r. From this we obtain Monte Carlo approximation of the mean values and standard deviations of the estimates. Two different trajectories of length 5 and are considered and ensuing results are shown in Table. The results are satisfactory and comparable for both methods with acceptable variances of the estimators. Although the estimation accuracy is better for longer trajectory, it is still acceptable even for the sample size of n = 5. Since we deal here with fairly dependent data, estimation accuracy will largely depend on the strength of the dependence in the data closer ρ x and ρ r are to one more data is needed to estimate these parameters. In both of the zero-crossing and autocorrelation methods, estimation of ρ x performs similarly. For lower values of ρ r, the autocorrelation method estimates ρ r accurately. For higher values of ρ x and ρ r, the zero-crossing method appears to be better. Variances for estimates of ρ r are quite comparable and in some cases somewhat smaller for the zero-crossing method but evidence is not compelling. In addition, the value of ν parameter has not any significant effect on the estimation of ρ x and ρ r but the estimate of ν is quite variable and appears to be slightly biased 9

11 Elevation [m] Distance [km] Figure 5: Top: The topography (elevation in [m]) of the road. Bottom: Graphical demonstration of defined hilliness characteristics. for the larger ν and ρ r. These are preliminary findings and more studies are needed to confirm these observations. 6 Modeling slopes in a mine Here the Laplace autoregressive model is used to describe a real road topography data set. In Figure it was shown that the long term distribution of encountered slopes is well described by means of a generalized Laplace distribution. It is rather unreasonable to expect that the complex dependence in the actual data can be described completely by our simple model. While, as mentioned before, it is possible to consider a more complicated Gaussian component, the goal here is to examine how well our two parameter dependence model captures the important hilliness characteristics encountered in the road profile, see Figure 5 (Top). 6. Hilliness characteristics In order to discuss properties of our random model for slopes variability, road topography and, in particular, variability of their hilliness, there is a need for a mathematical definition of some characteristics of the road topography. These definitions are represented graphically using a portion of hilliness data in Figure 5 (Bottom). A hill starts when the slope becomes positive and continues as long as one drives upwards to the crest. Then slope becomes negative and a hill continues until the slope becomes positive again and the next hill starts. Mathematically speaking, a hill is a part of a sequence of slopes Y k, say, between two consecutive upcrossings of zero level by Y k. The horizontal length of a hill is the distance between

12 Hill length L [m] Hill height H [m] Slope [%] Number of hills Number of hills Number of hills Figure 6: Comparison of amplitude spectra of hill lengths L (Left), hill heights H (Middle) and average slopes S (Right) encountered in the actual data (dashed lines) and in simulated ones. Simulated Y k process is defined by fitted parameters. The zero-crossing method is represented by thick lines and the autocorrelation method by the thin lines. two consecutive upcrossings. Here we split a hill in the part where road goes up (uphill) and the part when it goes down (downhill). The length of the two parts are denoted by L + and L, respectively. The steepness is described by averaging slopes. Namely, S + is the average slope on the uphill while S is the average slope on the downhill. Clearly, a hill height is H + =. S + L + on the uphill side and H =. S L on the downhill side (S is measured in %). We aim to find a model of the time series of slopes Y k which theoretically provides distribution of the hills characteristics. Since the model is symmetric, the model characteristics L +, S +, H + will have the same distribution as L, S and H, respectively. For simplicity of the presentation, in the following we will only consider characteristics L +, S +, H + and denote them by L, S, H, respectively. We shall demonstrate that the important hill characteristics; horizontal length L, average slope S, and vertical height H are reasonably well modeled by means of the introduced model. Distributions of hill characteristics are compared using the so-called amplitude spectrum that is defined as follows. Suppose that we have a sequence of positive quantities (amplitudes) z i >, i =,..., n. The observations have the cumulative distribution function (cdf) F Z (z). To illustrate the distribution of z i, instead of plotting the pairs (z, F Z (z)), one can alternatively use the amplitude spectrum which is a graph of (n ( F Z (z)), z), z. In a plot of amplitude spectra of two sequences, one can verify whether the signals have the same distribution of large to moderate amplitudes. This is often used in fatigue analysis to compare two sequences of cycles amplitudes, see e.g. []. 6. Hilliness characteristics using the zero-crossing method Variability of L - estimation of the parameter ρ x We begin with the horizontal length L that is completely determined by time series X k, i.e. a Gaussian AR() process having parameter ρ x. Since the records are measured every [m], the estimated intensity of hills is ˆf/ =.53 [m ] which gives E [L] = / ˆf = 94.9[m] and also, by (), ρ x =.79. We turn next to comparison of the variability of L i found in the measured signal and extracted from a simulation of Gaussian AR() processes of the same length as the measured signal. In Figure 6 (Left), the dashed line represents the amplitude spectrum of the observed horizontal lengths L i, i =,..., n. The amplitude spectra from simulations for the horizontal lengths of the hills each fit by the zero-crossing method have been averaged and presented using thick lines. We observe a quite good agreement between the model and the empirical amplitude spectrum for the zero-crossing method marked by the thick solid line. In this figure it is also marked by the thin line, the corresponding amplitude spectra for the autocorrelation fit to the data. The result shows that the autocorrelation fitting slightly overestimates sizes of the hill lengths.

13 Autocorrelation of Y.6.4. Autocorrelation of Y Lag in [m] Lag in [m] Figure 7: Comparison of the autocorrelations ˆρ y (k), ˆρ y(k), left, right plots respectively, fitted to the data presented in Figure. The thick dashed lines are the empirical autocorrelations. The thin solid lines are the theoretical autocorrelations of the introduced model with parameters ρ x =.85, ρ r =.679 fitted using the autocorrelation method, while in thick solid lines, the fit obtained from the zero-crossing method yields parameters ρ x =.789, ρ r =.839 and is presented. Variability of S and H - estimation of the parameter ρ r Since the correlation ρ x is already estimated in the previous section, it remains to fit ρ r. For this, we use relation () in which ν has be replaced by its estimated ˆν =. obtained from the empirical kurtosis. We apply (7) for this purpose, as described in Section 5. In our particular example, we have chosen K = 3, which corresponds to using autocorrelations at [m], 4[m], and 6[m] a reasonable choice of lags for the topography records. This has yielded ˆρ r =.84. In Figure 7, we present the sample estimate of ρ y (k) versus the one obtained from the model using the fitted values of ρ r and ρ x. The fits are quite reasonable but our model does not capture all the dependencies observed in the data. This could be remedied by considering more general correlation (spectra) for the Gaussian component. This approach, however, is not explored in this paper as our simple model seems to capture well the essential topographical features of the data. To seek confirmation of the adequacy of the fitted gamma autoregressive variance AR() model, we present averages of the vertical heights of hills H and the extracted slopes S based on simulations. The resulting amplitude spectra are shown in Figure 6 (Middle) and (Right) using a thin line. The thick lines are amplitude spectra found in the measured signal. We observe that the model captures the distribution of the height of hills and average slope. In fact, as expected, the zero-crossing model performs better than the autoregressive model in retrieving fairly accurate amplitude spectra. Nevertheless, both the fits are fairly accurate as far as to capture the fundamental hill characteristics. 6.3 Hilliness characteristics using the autocorrelation method As described in Section 5., we can alternatively use the one-step correlation formula for Y k to relate ρ r with ρ x () and then use () to fit ρ x (k), which might be very usefull especially for more general models of ρ x (k). The solution ˆρ x to this equation has to be found numerically due to the involvement of non-linear function of ρ r in (8). Using the autocorrelation method, we have obtained ˆρ x =.85 and ˆρ r =.68. As shown in Figure 7, autocorrelation functions are estimated in a satisfactory manner by both the methods. 7 Simulating from the model For convenience of the reader, we present a MATLAB script to simulate the discussed model and use it to simulate samples from the model. >> N=65; >> sy=4.8; nu=.; >> rho_x=.789; rho_r=.839; >> rar=zeros(n,); xar=zeros(n,); >> riid=nu*wgamrnd(/nu,,n,); xiid=wnormrnd(,,n,);

14 Slope topography 5 Slope [%] Distance [km] Simulated slope topography 5 Slope [%] Distance [km] 6 Simulated autoregresive Gamma variance 5 4 R Distance [km] Figure 8: (Top): Observed road slope topography. (Middle): Simulated road slope topography. (Bottom): Corresponding simulated autoregressive gamma variance. The Laplace model has the following parameters; the variance σ y = 4.8, the distributional parameter ν =. and the correlations are ρ x =.789 and ρ r =.839. >> rar()=riid(); xar()=xiid(); >> for i=:n >> m=rho_r/(-rho_r)/nu*rar(i-); >> NPois=poissrnd(m); >> if NPois>, W=wexprnd(,NPois,); else W=; end >> kj=sum(w)/m; >> rar(i)=rho_r*kj*rar(i-)+(-rho_r)*riid(i); >> xar(i)=rho_x*xar(i-)+sqrt(-rho_x^)*xiid(i); >> end >> y=sqrt(sy)*xar.*(sqrt(rar)); >> y=y-mean(y); >> plot(:n,cumsum(y)) 3

15 6 Observed and simulated elevation topography 4 Elevation [m] Distance [km] Figure 9: Three samples of simulated elevations (thin solid lines) are compared with the observed road elevation topography (solid thick line). The Laplace model has the following parameters; the variance σy = 4.8, the distributional parameter ν =. and the correlations ρ x =.789, ρ r =.839. In the program the variance of Y k, σ y = 4.8, the distributional parameter ν =. while the correlations ρ x =.793, ρ r =.839. Using the MATLAB code, the road slope topography and the corresponding autoregressive gamma variance are simulated and plotted in Figure 8 together with the observed slope topography of a [km] section. The samples have the same number of points as the measured slopes in that section, i.e. N = 6. Using the mentioned code, three random elevation topographies are evaluated and plotted in Figure 9. The simulated topographies are compared with the observed elevation topography presented using a thick solid line. Here N = 6 which corresponds to the size of the entire data set. The changes in variability of the observed slope topography of the road are captured fairly well in the simulated slopes and also the observed elevation topography is well emulated by the model. 8 Conclusions In the paper we discuss an extension of the classical Gaussian autoregressive models to account for non- Gaussian distribution of the records as well as to provide a more flexible representation of topography in the data of a road. The model has two components: the first one is given by a classical Gaussian AR() time series while the second one represents the randomly varying variance given by an autoregressive gamma process. The dependence in the model is governed by two autoregressive parameters: ρ x for the Gaussian AR() part and ρ r for the autoregressive gamma part. The distributional parameters, location, scale and shape are obtained by fitting sample mean, variance and kurtosis to their sample equivalents. We consider two estimation methods to fit the autoregressive parameters. In the first one, the parameter ρ x is a function of the expected length of the hills and can be simply fit using the empirical distribution of the length. The parameter ρ r is harder to estimate since the distributions of the height of the hills and their steepness are both influenced by its value. We used a simple relation between ρ r, ρ x and the autocorrelation of the squared values of the slopes to obtain an estimate of ρ r. In the second method, we have used the one-lag correlation of the slope values instead of the expected length of the hills. We illustrate that both the methods perform quite well for simulated data. We also fit the model to an example of real-life road topography data. It is shown that the model satisfactory retrieves the distributions of the length, the average slope, and the height of the hills. The model uses only the order-one autoregressive dependence and thus is not capable of capturing longer lag dependencies that are also present in the data. This can be remedied by an extension that involves the Gaussian component with a more complex dependence structure. This direction of research will be 4

16 pursued in future. Appendix Autocorrelation function of Y k and Y k The goal of this appendix is to present the argument for the theses in Proposition. Generally, without assuming anything except than independence between R k and X k, ρ y (k) is given by ρ y (k) = ρ r(k)ρ x (k) + ρ r (k)cv x + cv x + cv r + ρ x (k)cv r where ρ r (k) = corr (R, R k ), ρ x (k) = corr ( X, X k), and cvx, cv r are coefficients of variation for X and R, respectively. Similarly, the correlation of Y can be written as ρ y (k) = ρ r/(k) + cv r / + cv r / ρ x (k). If we add the assumption that X k is a Gaussian process, then we have Cov ( X, X k) = [ Cov (X, X k ) ],, which leads us to ρ y (k) = ρ r(k)ρ x(k) + ρ r (k)/ + ρ x(k)cvr 3/ + cvr. Further, by assuming that R k is autoregressive gamma and utilizing the formula for the moments of a gamma distribution we get cv r = ν and ρ r (k) = ρ k r, where ρ r is set to ρ r () which leads to ρ y (k) = νρ x(k)ρ k r + νρ k r + ρ x(k). () 3ν + Similarly, since where we obtain cv r / = a ν, a ν = Γ( ν + )Γ( ν ) Γ ( ν + ), ρ y (k) = [ρ r /(k) + a ν ( ρ r /(k))] ρ x (k). For our special AR() example, it is equal to Acknowledgments ρ y (k) = (ρk x + )(ν ρ k r + ). (3) 3ν + The authors are thankful to Volvo Construction Equipment for supplying data. The first and third authors were support by The Swedish Energy Agency. The research of the second and the fourth authors was supported by the Riksbankens Jubileumsfond Grant Dnr: P3-4: and the Swedish Research Council Grant Dnr:

17 References [] A. Baxevani, K. Podgórski, and J. Wegener. Sample path asymmetries in non-gaussian random processes. Scandinavian Journal of Statistics, pages 3, 4. [] K. Bogsjö. Evaluation of stochastic models of parallel road tracks. Probabilistic Engineering Mechanics, :36 37, 7. [3] K. Bogsjö. Road Profile Statistics Relevant for Vehicle Fatigue. PhD thesis, PhD thesis, Mathematical Statistics, Lund University, 7. [4] K. Bogsjö, K. Podgorski, and I. Rychlik. Models for road surface roughness. Vehicle System Dynamics, 5:75 747,. [5] B. Bruscella, V. Rouillard, and M. Sek. Analysis of road surfaces profiles. Journal of Transportation Engineering, 5:55 59, 999. [6] D. Charles. Derivation of environment descriptions and test severities from measured road transportation data. Journal of the IES, 36:37 4, 993. [7] C. J. Dodds and J. D. Robson. The description of road surface roughness. Journal of Sound and Vibration, 3:75 83, 973. [8] P. Johannesson, K. Podgórski, and I. Rychlik. Modelling roughness of road profiles on parallel tracks using roughness indicators. Technical Report 4, MV Chalmers Preprint 4:4, 4. [9] P. Johannesson and I. Rychlik. Modelling of road profiles using roughness indicators. International Journal of Vehicle Design, pages , 4. [] P. Johannesson and M. Speckert, editors. Guide to Load Analysis for Durability in Vehicle Engineering. Wiley-Chichester, 3. [] S. Kotz, T.J. Kozubowski, and K. Podgórski. The Laplace Distribution and Generalizations: A Revisit with Applications to Communications, Economics, Engineering and Finance. Birkhaüser, Boston, New York,. [] T. J. Kozubowski and K. Podgórski. Skew Laplace distributions II. Divisibility properties and extensions to stochastic processes. Math. Scientist, 33:35 48, 8. [3] M. Kvanström, K. Podgórski, and I. Rychlik. Laplace moving average model for multi-axial responses in fatigue analysis of a cultivator. Probabilistic Engineering Mechanics, 34: 5, 3. [4] P. Mucka. Road waviness and the dynamic tyre force. Int. J. Vehicle Design, 36:6 3, 4. [5] V. Rouillard. Decomposing pavement surface profiles into a Gaussian sequence. Int. J. Vehicle Systems Modelling and Testing, 4:88 35, 9. [6] C. H. Sim. First-order autoregressive models for gamma and exponential processes. Journal of Applied Probability, 7:35 33, 97. [7] L. Sun, Z. Zhang, and J. Ruth. Modeling indirect statistics of surface roughness. Journal of Transportation Engineering, 7:5,. 6

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