Simultaneous Input and State Estimation of Linear Discrete-Time Stochastic Systems with Input Aggregate Information

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1 21 IEEE 4th Annual Conference on Decision and Control (CDC) December 1-18, 21. Osaa, Japan Simultaneous Input and State Estimation of Linear Discrete-Time Stochastic Systems with Input Aggregate Information Sze Zheng Yong a Minghui Zhu b Emilio Frazzoli a Abstract In this paper, we present filtering algorithms for simultaneous input and state estimation of linear discretetime stochastic systems when the unnown inputs are partially nown, i.e., when some aggregate information of the unnown inputs is available as linear equality or inequality constraints. The stability and optimality properties of the filters are presented and proven using two complementary perspectives. Specifically, we confirm the intuition that the partial input information improves the performance of the filters when a linear input equality constraint is given. On the other hand, given a linear inequality constraint, we show that the estimate error covariance is decreased but the estimates may be biased. I. INTRODUCTION The estimation problem for stochastic systems with unnown inputs has applications found across a wide range of disciplines. The unnown disturbance inputs often cannot be modeled by a zero-mean, Gaussian white noise, but are typically not completely unnown, as they may satisfy conservation laws [1] or are bounded by physical laws. For instance, autonomous vehicles do not have nowledge of the control inputs of other vehicles [2] but these inputs are limited by the maximum engine power. Other application examples include real-time estimation of mean areal precipitation during a storm [3], fault detection and diagnosis [4] as well as population and traffic estimation [1], [], []. Literature review. In the extreme scenario where all inputs are observed, the seminal Kalman filter produces statistically optimal estimates of the underlying system states. For the other extreme where the inputs are completely unnown, numerous filters have been recently developed that find minimum-variance unbiased (MVU) estimates of the states only (see, e.g., [3], [7], [8]) or that simultaneously obtain MVU estimates of both states and inputs (see, e.g., [9] [12]). Relatively few filters have been developed for the case when the inputs are partially nown, although this is a nown problem for a long time. Initial research has assumed that these unnown inputs are fixed biases [13] or have nown dynamics [], [14]. More recently, the unnown inputs are assumed to be nown at an aggregate level [1], [1] (with equality), which was shown to be suitable for describing conservation laws in a traffic study [] and aggregated statistics of a population estimation study []. However, the proposed filters in [1], [1] assume that the system has no direct feehrough and only estimate states, although the problem of estimating the partially nown inputs is often as important as state estimation. On another hand, to our best nowledge, no filters have been developed for a S.Z Yong and E. Frazzoli are with the Laboratory for Information and Decision Systems, Massachusetts Institute of Technology, Cambridge, MA, USA ( szyong@mit.edu, frazzoli@mit.edu). b M. Zhu is with the Department of Electrical Engineering, Pennsylvania State University, University Par, PA, USA ( muz1@psu.edu). the scenario when the aggregate input is nown to satisfy a linear inequality, which, for instance, is the case when the unnown inputs are nown to be bounded below (e.g., nonnegative) or above (e.g., bounded L 1 - and L -norms). Thus, the problem of simultaneously estimating states and inputs in an unbiased minimum-variance manner for systems with linear input aggregate information and without restrictions on the direct feehrough matrix remains open. A related set of relevant literature pertains to constrained state estimation, i.e., Kalman filtering with linear or inequality constraints [1], [17]. Of particular interest to the filter development in this paper is the projection method, which exhibits several desirable properties and for which the constrained estimation problem can be either solved in closed form or with a quadratic program. Contribution. We introduce filtering algorithms for simultaneously estimating both states and unnown inputs when input aggregate information is available in the form of linear equality and inequality constraints. With the input aggregate equality constraint, we can transform the problem via a substitution method into an equivalent problem with no constraints. Hence, the nice properties of the input and state filter developed in an earlier wor [12] directly apply, e.g., optimality in a minimum variance unbiased sense, stability and convergence to steady-state. Furthermore, via a detour of using a projection method [1] that projects the unconstrained input estimates onto the constraint manifold, we can further show that with the input aggregate information, the mean-squared error of the estimates is decreased. In contrast, when the input aggregate information is given in the form of a linear inequality, we use the projection method to project the unconstrained input estimates onto the polyhedron described by the linear constraints instead. We show that the error covariance of the input estimates is decreased, but the estimates may become biased, although the bias is imperceptible in our simulation examples. Notation. R n denotes the n-dimensional Euclidean space and C the field of complex numbers. For a vector, v R n, the expectation and L 1 -, L 2 - and L -norms are denoted by E[v], v 1, v 2 and v. The transpose, inverse, Moore- Penrose pseudoinverse, trace and ran of M R p q are M, M 1, M, tr(m) and r(m). For a symmetric matrix S, S (S ) indicates S is positive (semi-)definite. II. PROBLEM STATEMENT AND MOTIVATION Consider the linear time-varying discrete-time system x +1 = A x + B u + G d + w y = C x + D u + H d + v (1) where x R n is the state vector at time, u R m is a nown input vector, d R p is an unnown input vector, /1/$ IEEE 41

2 and y R l is the measurement vector. The process noise w R n and the measurement noise v R l are assumed to be mutually uncorrelated, zero-mean, white random signals with nown covariance matrices, Q = E[w w ] and R = E[v v ], respectively. Without loss of generality, we assume that n l 1, l p and m, and that the current time variable r is strictly nonnegative. x is assumed to be independent of v and w for all. The matrices A, B, C, D, G and H are nown, and no assumption about the ran of H is made. We also assume that max (r[g H ]) = p (otherwise, we can retain the linearly independent columns and the remaining inputs still affect the system in the same way). In our previous wor [11], [12], we assumed that d for all is completely unnown. In this paper, we consider the case when partial input information is available, i.e., we are given input aggregate information in these linear forms: 1) Linear equality constraint: R d = ρ, (2) where ρ R r and R R r p are deterministic and nown. Furthermore, we assume that R has full row ran, i.e., ran(r ) = r and r p (otherwise, redundant constraints can be removed). This form of aggregate information is found in various contexts (cf. [1]) such as conservation laws, nown weighted averages, aggregated statistics, etc. Some concrete examples are when net migration arrivals (input variables) are only nown at a national (aggregated) level in the estimation problem of Australian state populations [] or when the net gain of lane-changing vehicles (unnown inputs) aggregated across all the lanes is equal to zero in a traffic densities study []. 2) Linear inequality constraint: R d ρ, (3) where ρ R r and R R r p are deterministic and nown (r is generally greater than p). This partial information form allows for the incorporation of input bounds (including L 1 -norm and L -norm) since unnown disturbance inputs in most practical problems are bounded. For instance, the unnown inputs of other drivers [2] are bounded by the vehicles engine power, whereas the fault signals in fault detection and diagnosis [4] are oftentimes bounded. In the context of data/signal injection attacs [18], [19] in cyberphysical systems, it may be reasonable to assume that the attacer has limited resources. The estimator design problem can thus be stated as: Given a linear discrete-time stochastic system (1) with input aggregate information as a linear equality constraint (2) and/or a linear inequality constraint (3), design an optimal and stable filter that simultaneously estimates system states and unnown inputs. III. FILTERING WITH PARTIAL INPUT INFORMATION Having motivated the problem at hand, we now proceed to present filtering algorithms for the case when input aggregate information is available, either as an equality or an inequality constraint. For the scenario when input equality information Algorithm 1 ULISE algorithm (for unconstrained/equality constrained unnown inputs) 1: Initialize: P x = P x = (C2,R 1 2, C2,) 1 ; ˆx = E[x ] = P C x 2,R 1 2, (z2, D2,u); Â = A G1,Σ 1 C1,; ˆQ = G 1,Σ 1 R1,Σ 1 G 1, + Q ; ˆ, = Σ 1 (z1, C1,ˆx D 1,u ); P1, d = Σ 1 (C1,P C x 1, + R 1,)Σ 1 ; 2: for = 1 to N do Estimation of d 2, 1 and d 1 3: Â 1 = A 1 G 1, 1 M 1, 1 C 1, 1 ; 4: ˆQ 1 = G 1, 1 M 1, 1 R 1, 1 M1, 1G 1, 1 + Q 1 ; : P = Â 1P 1 1Â x 1 + ˆQ 1 ; : R2, = C 2, P C2, + R 2, ; 7: P2, 1 d = (G 2, 1C2, 1 R 2, C 2,G 2, 1 ) 1 ; 8: 9: M 2, = P2, 1G d 2, 1C2, 1 R 2, ; ˆx 1 = A 1ˆx B 1 u 1 + G 1, 1 ˆdu 1, 1 ; 1: ˆ, 1 = M 2, (z 2, C 2, ˆx 1 D 2, u ); 11: ˆd 1 = V 1, 1 ˆ, 1 + V 2, 1 ˆ, 1 ; 12: P12, 1 d = M 1, 1 C 1, 1 P 1 1A x 1C2,M 2, P d 1, 1G 1, 1C 2,M 2,; 13: P 1 d P d = V 1, 1 P12, 1 d 1 P12, 1 d P d 2, 1 V 1; Time update 14: ˆx = ˆx 1 + G 2, 1 ˆ, 1 ; 1: P x = G 2, 1 M 2, R 2, M 2,G 2, +(I G 2, 1 M 2, C 2, ) P (I G 2, 1 M 2, C 2, ) ; 1: R 2, = C 2, P x C 2, + R 2, C 2, G 2, 1 M 2, R 2, R 2, M 2,G 2, 1C 2, ; Measurement update 17: L = (P x C2, G 2, 1 M 2, R 2, ) R 2, ; 18: ˆx = ˆx + L (z 2, C 2, ˆx D 2, u ); 19: P x = (I L C 2, )G 2, 1 M 2, R 2, L + L R 2, M2,G 2, 1(I L C 2, ) +(I L C 2, )P (I x L C 2, ) + L R 2, L ; Estimation of d 1, 2: R1, = C 1, P C x 1, + R 1, ; M 1, = Σ 1 ; 21: P1, d = M 1, R1, M 1, ; 22: ˆ, = M 1, (z 1, C 1, ˆx D 1, u ); 23: end for (2) is accessible, we consider two equivalent views of the input aggregate information: (i) as information that can be included or substituted into the system equations, or (ii) as a constraint that the input estimate needs to satisfy and also a manifold onto which the estimate can be projected. Fortunately, these two complementary perspectives provide the possibility to study different properties of the filter. Note, however, that the second perspective is solely given for analysis purposes, for reasons that we shall provide at the end of Section III-A. In addition, taing the perspective of input inequality information (3) as a constraint provides a means to project the input estimates onto the polyhedron described by the linear constraints. However, less can be said about the optimality of the filter in this case. The error covariance of the input estimates decreases with the projection, but the estimate may be biased (though imperceptible in simulations); thus, improvements of the filter remain a subject of future wor. In the following, we denote the error covariances of the constrained (i.e., with input aggregate information) propagated and updated state estimates as P x := Cov( x ) and 42

3 P x := Cov( x ), the error covariances of the constrained input estimate as P d 1 := Cov( d 1 ), and the unconstrained input estimate (i.e., when ignoring the input aggregate information) as 1 := Cov( d u 1 ). A. Filtering with Input Aggregate Equality Information 1) Input Equality as Information (via substitution): Viewing the given input aggregate equality (2) as information, we incorporate this partial information via substitution into (1). Note that if r < p, there exists an infinite number of solutions to (2). Thus, we first obtain all solutions to (2) as d = R ρ + (I R R ) d = R ρ + U 1, Ξ δ, (4) where the latter term represents all possible signals in the null space of R and we denote with R := R (R R ) 1 the Moore-Penrose pseudoinverse of R. We have also carried out a singular value decomposition of the matrix I R R I R R := [ U 1, U 2, ] [ Ξ ] V 1, = U 1, Ξ V 1,, V 2, and defined δ := V d 1, R p with p = p r. It can be verified that (4) satisfies (2) for all d R m. Then, substituting (4) into the original system (1) and rearranging, we obtain an equivalent system given by x +1 = A x + B u + G δ + w y = C x + D u + H δ () + v where B := [ B G R ], D := [ D H R ], G := G U 1, Ξ, H := H U 1, Ξ and u =, u ρ which has the exact form as (1) except with matrices (A, B, C, D, G, H ) and with the noise statistics, Q and R, remaining unchanged. Thus, the ULISE algorithm (cf. Algorithm 1) can be directly applied and by extension, all the nice properties of ULISE such as (i) optimality in the unbiased and minimum variance sense, (ii) the global optimality over the class of all linear state and input estimators, and (iii) stability guarantees and convergence to steady-state (for LTI systems) also hold for the input equality constrained problem. For the sae of brevity, the reader is referred to [12] for proofs and a detailed discussion. Note that some components of δ may only be estimated with a delay [12], [2], which implies that some components of d may only be estimated with a delay too. As discussed in [12], [2], strong detectability is a ey property for the existence and stability of a simultaneous input and state filter for linear time-invariant systems. Naturally, we would expect that the strong detectability of the original system without input aggregate information (1) would imply that the equivalent system with partial input information () is also strongly detectable, whereas the converse may not be true. Thus, the next proposition formalizes the conditions for strong detectability of the equivalent linear time-invariant system with input aggregate information (), and provides a proof that confirms the intuition. Proposition 1 (Strong detectability (time-invariant)). Given an input aggregate equality constraint (2), the equivalent system () is strongly detectability if and only if zi A G I r = n + p, z C, z 1, () C H U 1 where p := p r(r) =: p r. Moreover, strong detectability of the original LTI system (1) (without input equality information) given by zi A G r = n + p z C, z 1, (7) C H is sufficient but not necessary for the strong detectability condition in (). Proof. From the necessary and sufficient condition for the LTI system (A, B, C, D, G, H ) given in [12], we have zi A GU1 Ξ zi A G I I r = r C HU 1 Ξ C H U 1 Ξ zi A G I = r = n + p, z C, z 1, C H U 1 where we [ used the fact that ] Ξ is invertible. Next, if (7) holds, zi A GU1 Ξ I then r = r = n + p, i.e., (7) C HU 1 Ξ U 1 implies (). The reverse is not necessarily true, as[ is shown ].1 1 with the following unstable LTI system A =, G =, C = and H =, in which (7) does not hold because of an invariant zero at {1.} but with input aggregate information (2) with R = [ 1 1 ], the strong detectability condition () holds. 2) Input Equality as Constraint (via projection): If we view the input aggregate information as a constraint, we can further prove that the estimates have better performance (smaller mean-squared errors in the estimates of states and inputs) than without the constraint (similar to state-constrained Kalman filtering [1]), in addition to the properties that are carried over from the properties of ULISE as previously discussed in Section III-A.1. To achieve this, we first show that a commonly employed projection method in [1] leads to estimates in the linear form, which have the desired properties. Then, from the global optimality of the ULISE algorithm over the class of linear estimators [12], we argue that the estimates computed via substitution in Section III-A.1 also have the same properties. The projection method in [1] is useful for projecting the unconstrained input estimates ˆd u onto the constraint surface. Thus, to obtain constrained input estimates, we solve the problem given by ˆd := arg min (δ ˆd u ) W (δ ˆd u ) s.t. R δ = ρ (8) δ where W is any symmetric positive definite weighting matrix (which we shall see is optimal when W = ( ) 1 ). This constrained optimization problem can be solved analytically, yielding ˆd = ˆd u J (R ˆdu ρ ), (9) where J := W 1 R (R W 1 R ) 1. Moreover, we can find the input estimate error, d := d ˆd, from (9) as d = d ˆd u + J (R ˆdu ρ R d + R d ) = (I J ) d u + J (R d ρ ), (1) where d u := d ˆd u and J := J R. Note that in this input equality case, R d ρ =, which further simplifies (1) to d = (I J ) d u, and this is the crux for the derivation of 43

4 estimate properties, which we state in the following without proof (interested readers are referred to [1, Theorems 1 3]). Proposition 2. Let the initial state estimate ˆx be unbiased. Then the input estimate is unbiased, i.e., E[ ˆd 1 d 1 ] =. Proposition 3. The constrained input estimate ˆd as given by (9) with W d,u = (P ) 1 has a smaller error covariance than the unconstrained input estimate ˆd u, i.e., P d. Proposition 4. Among all (symmetric positive definite) weighting matrices, W, the estimator of d that uses W := ( ) 1 has the smallest estimation error covariance P d, i.e., P d P d,w. A straightforward corollary follows: Corollary 1. The constrained and unconstrained input estimate errors also satisfy trace(p d d,u ) trace(p ) and trace(p d) trace(p d,w ) for any W ; and equivalently, E[ d 2 2] E[ d u 2 2] and E[ d 2 2] E[ d W 2 2] if ˆx is unbiased. Proof. By Proposition 3, P d. Since the trace of negative semidefinite matrices is non-positive, it follows that trace(p d ) trace(p d d,u ) trace(p ). The latter trace inequality holds similarly from Proposition 4. Finally, the mean-squared error inequalities follows from the equivalence: tr(p d) = tr(e[ d d ]) = E[ d 2 2]. Therefore, we observe from the above claims that W = ( ) 1 is optimal in that it minimizes the trace of P d and thus the mean-squared error of the input estimates. Next, we view d as the sum of ˆd and a zero-mean Gaussian noise term, d. Thus, any additional uncertainty in the input estimate (trace( ) trace(p d ) by Corollary 1) will increase the effective noise affecting the system. Intuitively, the larger uncertainty in the noise terms should not lead to a smaller uncertainty in the state estimates, i.e., we expect trace(p x,u ) trace(p x) and E[ xu 2 2] E[ x 2 2], which we observe to hold in simulation and can be explained in steady-state by the following proposition. Proposition (Steady-state covariance (time-invariant)). Suppose that the filters for the linear time-invariant systems with and without input aggregate equality information are stable. Then, the steady-state error covariance of the state estimates with the input aggregate equality information/constraint, P, x is smaller than without the input equality, P x,u, i.e., we have P x P x,u and equivalently, trace(p ) x trace(p x,u ) and E[ x 2 2] E[ x u 2 2]. Proof. As discussed above, with the view that d = ˆd + d, the time-invariant filtering problem is equivalent to x +1 = Ax + Bu + G ˆd + W w eff (11) z 2, = C 2 x + D 2 u + v 2,, where ˆd is nown, W = [ G I ], w eff = d w and only the z 2, component of the measurement is used to ensure the unbiasedness of the estimates (cf. [12]). Note that we denote with ˆd and ˆd u (and, d and d u ) the constrained and unconstrained input estimates (and input estimate errors). By Proposition 3, we have P d, which implies that = P d + P d for P d, which essentially means that the effective process noise, w eff, is larger when no equality input information is incorporated, i.e., Cov(w eff,u ) Cov(w eff ) =: Qeff,u Q eff = Q for some Q. By assumption, filters for the linear time-invariant systems with and without input aggregate equality information are stable and hence the steady-state effective process noise covariance Q eff,u = Q eff + Q exists with Q. Thus, with the same arguments in [21, pp ] (briefly summarized here due to space constraints), a linearity property of the filters holds in steady-state for any filter gain matrix K, i.e., P x = (I KC 2 )(AP A x + W Q eff W )(I KC 2 ) +KR 2 K P x = (I KC 2 )(A P A x + W Q W )(I KC 2 ) P x,u = (I KC 2 )(AP x,u +KR 2 K A + W Q eff,u W )(I KC 2 ) where the third equation is a sum of the first two; hence, P x,u = P x + P, x where P x. It follows that P x P x,u and equivalently, trace(p ) x trace(p x,u ) and E[ x 2 2] E[ x u 2 2] (same proof as in Corollary 1). It is noteworthy that if only the component z 2, R l p H with p H := r(h ) is used in the measurement update in ULISE, instead of z 2, Rl p H with p H := r(h ) p H (that results from the equivalent system ()), a slight loss in optimality may be expected. Moreover, it is unclear if the input estimates obtained with the projection method is equivalent to the one with the previous perspective of the input aggregate nowledge as information, which also minimizes the trace of the input error covariance matrix [12]. Nonetheless, since the ULISE algorithm is globally optimal over the class of linear input and state estimators [12], we now that the estimates using the substitution method in Section III-A.1 are no worse than the estimates that result from the projection method, whose estimates are in the linear form (cf. (9)). This means that Propositions 2,3,4, and Corollary 1 also hold for the estimates using the method in Section III-A.1. Hence, we recommend the use of ULISE algorithm (see Algorithm 1) with the equivalent system () (i.e., to use the substitution method in Section III-A.1) over the projection method in this section. The projection method is provided mainly for analysis purposes and also in preparation for the next section on filtering with input aggregate inequality information. B. Filtering with Input Aggregate Inequality Information For the case with input inequality information, the projection method in [1] can be applied to project the unconstrained input estimates ˆd u onto the constraint polyhedron, i.e., we now solve the problem given by ˆd := arg min(δ ˆd u ) W (δ ˆd u ) s.t. R δ ρ (12) δ where W is any symmetric positive definite weighting matrix. This constrained optimization problem is a quadratic optimization problem and can be efficiently solved using offthe-shelf software pacages. From the complementary slacness condition of the Karush-Kuhn-Tucer (KKT) conditions for optimality, we now that we can find the corresponding active constraints (for each i such that λ i > ) from the KKT multiplier vector, λ, that is typically returned by the 44

5 optimization routine. We denote by R and ρ the rows of R and elements of ρ corresponding to the active constraints; thus, the constrained input estimate ˆd satisfies R ˆd = ρ. Similar to (9) and (1), we have in this case ˆd = ˆd u J ( R ˆdu ρ ) d = (I J ) d u + J ( R (13) d ρ ), where J := W 1 R ( R W 1 R ) 1 and J := J R. If J ( R d ρ ) = (implicitly assumed in [17]), then we have d = (I J ) d u and thus, Propositions 2, 3 and 4 would directly follow. However, we observed in our simulations that this condition does not generally hold. In this case, we show in the following that the estimates may no longer be unbiased, although the error covariance of the constrained estimates is still lower than that of the unconstrained estimates when d is deterministic and W = ( ) 1. Proposition. Suppose J ( R d ρ ) but E[ d u ] = (a property of ULISE [12]). Then, the inequality constrained input estimate ˆd is biased, i.e., E[ d ] = J ( R d ρ ). Proof. Although ˆd u is unbiased [12], since J ( R d ρ ), it follows from (13) that input estimate is biased. Proposition 7. Suppose d is deterministic. Then, the constrained input estimate ˆd as given by (12) with W = ( ) 1 has a smaller error covariance than the unconstrained input estimate ˆd u, i.e., P d, and equivalently, trace(p d d,u ) trace(p )). Proof. Since d is deterministic, from (13), we have P d := Cov( d ) = (I J )Cov( d u )(I J ) = (I J ) (I J ) = J J J + J =, where the final equality is obtained because it J (14) can be verified from the definition of J that J = J J with W = ( ) 1. Furthermore, J = R ( R R ) 1 R and thus, trace(j ). It then follows that P d and trace(p d d,u ) trace(p ). Proposition 8. Suppose d is deterministic. Among all (symmetric positive definite) weighting matrices, W 1, the estimator of d that uses W 1 = ( 1 ) 1 has the smallest estimation error covariance. Proof. Proceeding from (14), the proof is identical to [1, Theorem 3]. Next, with the inequality constrained input estimate ˆd 1 given in (12), we proceed as in the ULISE algorithm (cf. Algorithm 1) with the time update and measurement update steps to find the propagated and updated state estimates: ˆx = A 1ˆx B 1 u 1 + G 1 ˆd 1 (1) ˆx = ˆx + L (z 2, C 2, ˆx D 2,u ) (1) where the filter gain L can be chosen as in [12] to minimize the state estimate error covariance trace(p x ): L = (P x C 2, Ĝ2, 1M 2, R 2, ) R 2,, (17) where we defined Ĝ2, 1 := G 1 (I J 1 )V 2, 1, R 2, := C 2,P x C 2, + R 2, C 2, Ĝ 2, 1 M 2, R 2, R 2, M 2,Ĝ 2, 1 C 2, and the propagated state estimate error covariance is given by P x = [ P A 1 G 1 1 x P xd ] 1 A Q 1 P 1 xd P d 1 G 1 Ĝ2, 1M 2, C 2, Q 1 Q 1 C2, M2,Ĝ 2, 1, (18) while the resulting P x can be computed as P x = L R 2, L + (I L C 2, )Ĝ2, 1M 2, R 2, L + L R 2, M 2,Ĝ 2, 1 (I L C 2, ) +(I L C 2, )P x (I L C 2, ). (19) The derivation of the above equations is similar to [12] with the small difference that d is as given in (13), where its (unnown) deterministic term does not play a role in the computation of error cross-covariances if we assume that the state estimates remain unbiased. Since this may not hold in general because of the possible bias in the input estimates (Proposition ), the resulting filter may be suboptimal. Besides, due to the use of only z 2, (cf. discussion at the end of Section III-A.2), the optimality of this filter may further deteriorate. However, this loss of optimality is hardly noticeable in the simulations in Section IV-B, with this filter (summarized in Algorithm 2) still outperforming the case when the input inequality information is ignored. IV. ILLUSTRATIVE EXAMPLES A. Fault Identification We consider the benchmar problem of state estimation and fault identification [8], [12] when the system dynamics is plagued by faults, d, that influence the system dynamics through the input matrix G and the outputs through the feehrough matrix H as well as zero-mean Gaussian white noises. Moreover, we assume partial nowledge of the unnown inputs in the linear form Rd = ρ A = ; G = ; H = 1 ; Q = ; R = 1. 1 ; B = 1; C = I ; D = 1; R = 1 1. The unnown inputs used in this example are { 1, 7 d,1 =, otherwise { 1 ( 1), 1 8 d,2 = 7, otherwise { 3, 49, 49, d,3 = 3, 99, 99, 7 799, otherwise. Figure 1 shows a comparison of the input and state estimates with and without the input aggregate information (i.e., constrained by a linear equality and unconstrained, respectively). As expected, the estimates when given the 4

6 Algorithm 2 i-ulise algorithm (for inequality constrained unnown inputs) 1: Initialize: P x = P x = (C2,R 1 2, C2,) 1 ; ˆx = E[x ] = P C x 2,R 1 2, (z2, D2,u); Â = A G1,Σ 1 C1,; ˆQ = G 1,Σ 1 R1,Σ 1 G 1, + Q ; ˆdu 1, = Σ 1 (z1, C1,ˆx D 1,u ); P1, d = Σ 1 (C1,P C x 1, + R 1,)Σ 1 ; 2: for = 1 to N do Estimation of d 2, 1 and d 1 3: Â 1 = A 1 G 1, 1 M 1, 1 C 1, 1 ; 4: ˆQ 1 = G 1, 1 M 1, 1 R 1, 1 M1, 1G 1, 1 + Q 1 ; : P = Â 1P 1 1Â x 1 + ˆQ 1 ; : R2, = C 2, P C2, + R 2, ; 7: 2, 1 = (G 2, 1C2, 1 R 2, C 2,G 2, 1 ) 1 ; 8: M 2, = 2, 1 G 2, 1C2, 1 R 2, ; 9: ˆx 1 = A 1ˆx B 1 u 1 + G 1, 1 ˆdu 1, 1 ; 1: ˆdu 2, 1 = M 2, (z 2, C 2, ˆx 1 D 2, u ); 11: ˆdu 1 = V 1, 1 ˆdu 1, 1 + V 2, 1 ˆdu 2, 1 ; 12: 12, 1 = M 1, 1C 1, 1 P 1 1A x 1C2,M 2, 13: 1 = V 1 1, 1 [ G 1, 1C2,M ] 2,; 1, 1 12, 1 12, 1 2, 1 14: P xd,u 2, 1 = P x 1 1A 1C 2,M 2, 1: P xd,u 1 V 1; P xd,u 1, 1 G 1. 1C2,M 2,; = P xd,u 1, 1 V 1, 1 + P xd,u 2, 1 V 2, 1; 1: ˆd 1 = arg min δ (δ ˆd u 1) ( 1 ) 1 (δ ˆd u 1) subject to R 1 δ r 1 ; 17: R 1 and r 1 corresponding to active set; 18: J 1 = R 1 1[ R 1 R 1 1] 1 R 1 ; 19: Ĝ 2, 1 = G 1 (I J 1 )V 2, 1 ; 2: P 1 d = (I J 1 ) 1 (I J 1) ; 21: P 1 xd = P xd,u 1 (I J 1) ; Time update 22: ˆx = A 1ˆx B 1 u 1 + G 1 ˆd 1 ; 23: P x = P x 1 1 P 1 xd A A 1 G 1 1 P 1 xd P d 1 G 1 +Q 1 Ĝ2, 1M 2, C 2, Q 1 Q 1 C 2,M 2,Ĝ 2, 1; 24: R 2, = C 2, P x C 2, + R 2, C 2, Ĝ 2, 1 M 2, R 2, R 2, M 2,Ĝ 2, 1C 2, ; Measurement update 2: L = (P x C2, Ĝ2, 1M 2, R 2, ) R 2, ; 2: ˆx = ˆx + L (z 2, C 2, ˆx D 2, u ); 27: P x = (I L C 2, )Ĝ2, 1M 2, R 2, L + L R 2, M 2,Ĝ 2, 1(I L C 2, ) +(I L C 2, )P (I x L C 2, ) + L R 2, L ; Estimation of d 1, 28: R1, = C 1, P C x 1, + R 1, ; 29: M 1, = Σ 1 ; 3: 1, = M R 1, 1, M 1, ; 31: ˆdu 1, = M 1, (z 1, C 1, ˆx D 1, u ); 32: end for input aggregate information are closer to the true states and inputs and have lower error covariances (cf. Figure 2). B. 3-Area Power System We consider a power system [19], [22] with three interconnected control areas, each consisting of generators and loads, with tie-lines providing interconnections between areas. A simplified model of the control areas and the tie-lines is given by (cf. parameter definitions in [22, Chap. 1], [23]): x x1 ˆx unc 1 ˆx Time, 1 x3.. x3 ˆx unc 3 ˆx3 x2 4 2 x2 ˆx unc 2 ˆx Time, 1 x4 ˆx unc 4 ˆx4. x Time, Time, x. x ˆx unc ˆx ˆdunc 1 ˆ Time, Time, 2 1 ˆdunc 2 ˆ 2 ˆdunc 3 ˆ Time, Time, Fig. 1: Unconstrained ( unc ) and constrained estimates of states x 1, x 2, x 3, x 4, x, and unnown inputs d 1, d 2 and d 3 in Example IV-A. trace(p x ) 1 1 un 1 trace(p d ) Time, un Time, Fig. 2: Traces of unconstrained and constrained estimate error covariance of states, trace(p x ), and unnown inputs, trace(p d ) in Example IV-A. Control area i: (i {1, 2, 3}) d ω i + Di ωi P mech i j i + d P mechi + P mech i T CHi d P vi Tie-line power flow, P ij d P ij tie tie P ij tie Pv i T CHi =, = P L i, + Pv i T Gi + ωi R f i T G i, between areas i and j: = P ref i T Gi ; (2) = T ij ( ω i ω j ), P ji ij tie = Ptie, (21) where ω i, P mechi and P v i represent deviations of the angular frequency, mechanical power and steam-valve position from their nominal operating values. We assume that all states are measured (i.e., C q = I) where the system is affected by additive zero mean Gaussian white process and measurement noises with nown covariances Q = 1 2 diag(1, 1., 2, 1.2, 2., 1.4,.3, 2.11, 3,.2,.9, 1.8) and R = 1 2 diag(2.1,., 2.2,.2, 1.9, 1.4, 1.3, 1.1, 2.3, 1.2,.3, 1.8). Moreover, the measurements of ω i are injected with adversarial additive signals d =, d,1 d,2 d,3 where the attacer is limited in that d 1, or equivalently, d. The system parameters used in this example are D 1 = 3, R f 1 =.3, M a 1 = 4, T CH1 =, T G1 = 4, D 2 =.27, R f 2 =.7, M a 2 = 4, T CH2 = 1, T G2 = 2, D 3 = 2, R f 3 =.4, M a 3 = 3, T CH3 = 2, T G3 = 1, T 12 = 2.4, T 23 = 1., T 31 = 2.. 4

7 ω ω ω1 ω3 ˆω unc 1 ˆω1 ˆω unc 3 ˆω3 ˆdunc 2 ω ˆ 1 1 ω2 ˆω unc 2 ˆω2 ˆdunc 1 ˆdunc 3 Fig. 3: Unconstrained ( unc ) and constrained state and attac magnitude estimates in Example IV-B. trace(p x ) 1 1 un Time, trace(p d ) 1 ˆ ˆ un Time, Fig. 4: Traces of unconstrained and constrained estimate error covariance of states, trace(p x ), and unnown inputs, trace(p d ) in Example IV-B. A comparison of the input and state estimates with and without the linear inequality constraint is shown in Figure 3. We observe little or no improvement to the state estimation, presumably because the attac signals only affect the measurements and not the system dynamics. However, we see that the input estimates are closer to the true inputs and the predicted bias (see Proposition ) appears insignificant. This can probably be explained by noticing that the bias is introduced only when R ˆd = ρ but R d ρ. That R ˆd = ρ suggests that R d ρ and thus the bias in (13) is liely to be small. We also observe in Figure 4 that the error covariance of the input is decreased only when the estimates are projected onto the boundary of the polyhedron described by linear inequality constraints, as expected. This also suggests that linear inequality constraints are less informative than equality constraints. V. CONCLUSION We presented filtering algorithms for simultaneous input and state estimation of linear discrete-time stochastic systems for systems in which the inputs are not completely nown but some aggregate information of the unnown inputs is available, either as linear equality or inequality constraints. Using two complementary views of the partial input information, we study the properties of the filters including the conditions for stability and optimality in the minimumvariance unbiased sense. We show that the estimate error covariance of the filters decreases when input aggregate information is available in either form. Moreover, given an equality constraint, the estimates remain unbiased but when given as an inequality constraint, the estimates may be biased although the bias is imperceptible in our simulation example. ACKNOWLEDGMENTS This wor was supported by the National Science Foundation, grant # M. Zhu is partially supported by ARO W911NF (MURI) and NSF CNS-14. REFERENCES [1] B. Li. State estimation with partially observed inputs: A unified Kalman filtering approach. Automatica, 49(3):81 82, 213. [2] S.Z. Yong, M. Zhu, and E. Frazzoli. Generalized innovation and inference algorithms for hidden mode switched linear stochastic systems with unnown inputs. In IEEE Conference on Decision and Control (CDC), pages , December 214. [3] P. K. Kitanidis. Unbiased minimum-variance linear state estimation. Automatica, 23():77 778, November [4] R. Patton, R. Clar, and P.M. Fran. Fault diagnosis in dynamic systems: theory and applications. Prentice-Hall international series in systems and control engineering. 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