Herded Gibbs Sampling
Authors: Luke Bornn, Yutian Chen, Nando de Freitas, Maya Baya, Jing Fang, Max Welling
ICLR 2013

Abstract: The Gibbs sampler is one of the most popular algorithms for inference in statistical models. In this paper, we introduce a herding variant of this algorithm, called herded Gibbs, that is entirely deterministic. We prove that herded Gibbs has an $O(1/T)$ convergence rate for models with independent variables and for fully connected probabilistic graphical models. Herded Gibbs is shown to outperform Gibbs in the tasks of image denoising with MRFs and named entity recognition with CRFs. However, the convergence for herded Gibbs for sparsely connected probabilistic graphical models is still an open problem.

Herded Gibbs Sampling
Luke Bornn
Harvard University
bornn@stat.harvard.edu
Yutian Chen
UC Irvine
yutian.chen@uci.edu
Nando de Freitas
UBC
nando@cs.ubc.ca
Mareija Eskelin
UBC
mareija@cs.ubc.ca
Jing Fang
Facebook
jingf@cs.ubc.ca
Max Welling
University of Amsterdam
welling@ics.uci.edu
Abstract
The Gibbs sampler is one of the most popular algorithms for inference in statistical
models. In this paper, we introduce a herding variant of this algorithm, called
herded Gibbs, that is entirely deterministic. We prove that herded Gibbs has an
O(1/T) convergence rate for models with independent variables and for fully
connected probabilistic graphical models. Herded Gibbs is shown to outperform
Gibbs in the tasks of image denoising with MRFs and named entity recognition
with CRFs. However, the convergence for herded Gibbs for sparsely connected
probabilistic graphical models is still an open problem.
1 Introduction
Over the last 60 years, we have witnessed great progress in the design of randomized sampling
algorithms; see for example [16, 9, 3, 22] and the references therein. In contrast, the design of deter-
ministic algorithms for “sampling” from distributions is still in its inception [8, 13, 7, 20]. There are,
however, many important reasons for pursuing this line of attack on the problem. From a theoreti-
cal perspective, this is a well deﬁned mathematical challenge whose solution might have important
consequences. It also brings us closer to reconciling the fact that we typically use pseudo-random
number generators to run Monte Carlo algorithms on classical, V on Neumann architecture, comput-
ers. Moreover, the theory for some of the recently proposed deterministic sampling algorithms has
taught us that they can achieve O(1/T) convergence rates [8, 13], which are much faster than the
standard Monte Carlo rates of O(1/
√
T) for computing ergodic averages. From a practical perspec-
tive, the design of deterministic sampling algorithms creates an opportunity for researchers to apply
a great body of knowledge on optimization to the problem of sampling; see for example [4] for an
early example of this.
The domain of application of currently existing deterministic sampling algorithms is still very nar-
row. Importantly, there do not exist deterministic tools for sampling from unnormalized multivariate
probability distributions. This is very limiting because the problem of sampling from unnormalized
distributions is at the heart of the ﬁeld of Bayesian inference and the probabilistic programming
approach to artiﬁcial intelligence [17, 6, 18, 11]. At the same time, despite great progress in Monte
Carlo simulation, the celebrated Gibbs sampler continues to be one of the most widely-used algo-
rithms. For, example it is the inference engine behind popular statistics packages [17], several tools
for text analysis [21], and Boltzmann machines [2, 12]. The popularity of Gibbs stems from its
simplicity of implementation and the fact that it is a very generic algorithm.
Without any doubt, it would be remarkable if we could design generic deterministic Gibbs sam-
plers with fast (theoretical and empirical) rates of convergence. In this paper, we take steps toward
Authors are listed in alphabetical order.
1
arXiv:1301.4168v2  [cs.LG]  16 Mar 2013
achieving this goal by capitalizing on a recent idea for deterministic simulation known as herding.
Herding [24, 23, 10] is a deterministic procedure for generating samplesx ∈X⊆ Rn, such that the
empirical moments µ of the data are matched. The herding procedure, at iteration t, is as follows:
x(t) = arg max
x∈X
⟨w(t−1),φ(x)⟩
w(t) = w(t−1) + µ −φ(x(t)), (1)
where φ : X→H is a feature map (statistic) from Xto a Hilbert space Hwith inner product ⟨·,·⟩,
w ∈H is the vector of parameters, and µ ∈H is the moment vector (expected value of φ over
the data) that we want to match. If we choose normalized features by making ∥φ(x)∥constant for
all x, then the update to generate samples x(t) for t = 1,2,...,T in Equation 1 is equivalent to
minimizing the objective
J(x1,..., xT) =
µ −1
T
T∑
t=1
φ(x(t))

2
, (2)
where T may have no prior known value and ∥·∥ =
√
⟨·,·⟩is the naturally deﬁned norm based
upon the inner product of the space H[8, 4].
Herding can be used to produce samples from normalized probability distributions. This is done
as follows. Let µ denote a discrete, normalized probability distribution, with µi ∈ [0,1] and∑n
i=1 µi = 1. A natural feature in this case is the vector φ(x) that has all entries equal to zero,
except for the entry at the position indicated by x. For instance, if x = 2 and n = 5, we have
φ(x) = (0,1,0,0,0)T. Hence, ˆµ = T−1 ∑T
t=1 φ(x(t)) is an empirical estimate of the distribution.
In this case, one step of the herding algorithm involves ﬁnding the largest component of the weight
vector (i⋆ = arg maxi∈{1,2,...,n}w(t−1)
i ), setting x(t) = i⋆, ﬁxing the i⋆-entry of φ(x(t)) to one
and all other entries to zero, and updating the weight vector: w(t) = w(t−1) + (µ −φ(x(t))). The
output is a set of samples {x(1),...,x (T)}for which the empirical estimate ˆµ converges on the
target distribution µ as O(1/T).
The herding method, as described thus far, only applies to normalized distributions or to problems
where the objective is not to guarantee that the samples come from the right target, but to ensure that
some moments are matched. An interpretation of herding in terms of Bayesian quadrature has been
put forward recently by [14].
In this paper, we will show that it is possible to use herding to generate samples from more complex
unnormalized probability distributions. In particular, we introduce a deterministic variant of the
popular Gibbs sampling algorithm, which we refer to as herded Gibbs. While Gibbs relies on draw-
ing samples from the full-conditionals at random, herded Gibbs generates the samples by matching
the full-conditionals. That is, one simply applies herding to all the full-conditional distributions.
The experiments will demonstrate that the new algorithm outperforms Gibbs sampling and mean
ﬁeld methods in the domain of sampling from sparsely connected probabilistic graphical models,
such as grid-lattice Markov random ﬁelds (MRFs) for image denoising and conditional random
ﬁelds (CRFs) for natural language processing.
We advance the theory by proving that the deterministic Gibbs algorithm converges for distributions
of independent variables and fully-connected probabilistic graphical models. However, a proof es-
tablishing suitable conditions that ensure convergence of herded Gibbs sampling for sparsely con-
nected probabilistic graphical models is still unavailable.
2 Herded Gibbs Sampling
For a graph of discrete nodes G= (V,E), where the set of nodes are the random variables V =
{Xi}N
i=1, Xi ∈X, let πdenote the target distribution deﬁned on G.
Gibbs sampling is one of the most popular methods to draw samples fromπ. Gibbs alternates (either
systematically or randomly) the sampling of each variable Xi given XN(i) = xN(i), where iis the
index of the node, and N(i) denotes the neighbors of node i. That is, Gibbs generates each sample
from its full-conditional distribution p(Xi|xN(i)).
2
Algorithm 1 Herded Gibbs Sampling.
Input: T.
Step 1: Set t = 0. Initialize X(0) in the support of π and w(0)
i,xN(i)
in (π(Xi = 1|xN(i)) −
1,π(Xi = 1|xN(i))).
for t= 1→T do
Step 2: Pick a node iaccording to some policy. Denote w= w(t−1)
i,x(t−1)
N(i)
.
Step 3: If w> 0, set x(t)
i = 1, otherwise set x(t)
i = 0.
Step 4: Update weight w(t)
i,x(t)
N(i)
= w(t−1)
i,x(t−1)
N(i)
+ π(Xi = 1|x(t−1)
N(i) ) −x(t)
i
Step 5: Keep the values of all the other nodes x(t)
j = x(t−1)
j ,∀j ̸= iand all the other weights
w(t)
j,xN(j)
= w(t−1)
j,xN(j)
,∀j ̸= ior xN(j) ̸= x(t−1)
N(i) .
end for
Output: x(1),..., x(T)
Herded Gibbs replaces the sampling from full-conditionals with herding at the level of the full-
conditionals. That is, it alternates a process of matching the full-conditional distributions p(Xi =
xi|XN(i)). To do this, herded Gibbs deﬁnes a set of auxiliary weights {wi,xN(i) }for any value of
Xi = xi and XN(i) = xN(i). For ease of presentation, we assume the domain of Xi is binary,
X = {0,1}, and we use one weight for every i and assignment to the neighbors xN(i). Herded
Gibbs can be trivially generalized to the multivariate setting by employing weight vectors in R|X|
instead of scalars.
If the binary variableXihas four binary neighborsXN(i), we must maintain24 = 16weight vectors.
Only the weight vector corresponding to the current instantiation of the neighbors is updated, as
illustrated in Algorithm 1. The memory complexity of herded Gibbs is exponential in the maximum
node degree. Note the algorithm is a deterministic Markov process with state (X,W).
The initialization in step 1 guarantees that X(t) always remains in the support of π. For a deter-
ministic scan policy in step 2, we take the value of variables x(tN),t ∈N as a sample sequence.
Throughout the paper all experiments employ a ﬁxed variable traversal for sample generation. We
call one such traversal of the variables a sweep.
3 Analysis
As herded Gibbs sampling is a deterministic algorithm, there is no stationary probability distribution
of states. Instead, we examine the average of the sample states over time and hypothesize that it
converges to the joint distribution, our target distribution,π. To make the treatment precise, we need
the following deﬁnition:
Deﬁnition 1. For a graph of discrete nodes G= (V,E), where the set of nodes V = {Xi}N
i=1,
Xi ∈ X, P(τ)
T is the empirical estimate of the joint distribution obtained by averaging over T
samples acquired from G. P(τ)
T is derived from T samples, collected at the end of every sweep over
N variables, starting from the τth sweep:
P(τ)
T (X = x) = 1
T
τ+T−1∑
k=τ
I(X(kN) = x) (3)
Our goal is to prove that the limiting average sample distribution over time converges to the target
distribution π. Speciﬁcally, we want to show the following:
lim
T→∞
P(τ)
T (x) =π(x),∀τ ≥0 (4)
If this holds, we also want to know what the convergence rate is.
3
We begin the theoretical analysis with a graph of one binary variable. For this graph, there is only
one weight w. Denote π(X = 1)as πfor notational simplicity. The sequence of X is determined
by the dynamics of w(shown in Figure 1):
w(t) = w(t−1) + π−I(w(t−1) >0), X (t) =
{
1 if w(t−1) >0
0 otherwise (5)
Lemma 3 in the appendix shows that (π−1,π] is the invariant interval of the dynamics, and the
state X = 1is visited at a frequency close to πwith an error:
|P(τ)
T (X = 1)−π|≤ 1
T (6)
This is known as the fast moment matching property in [24, 23, 10]. We will show in the next two
theorems that the fast moment matching property also holds for two special types of graphs, with
proofs provided in the appendix.
Figure 1: Herding dynamics for a single variable.
In an empty graph, all the variables are independent of each other and herded Gibbs reduces to
running N one-variable chains in parallel. Denote the marginal distribution πi := π(Xi = 1).
Examples of failing convergence in the presence of rational ratios between theπis were observed in
[4]. There the need for further theoretical research on this matter was pointed out. The following
theorem provides formal conditions for convergence in the restricted domain of empty graphs.
Theorem 1. For an empty graph, when herded Gibbs has a ﬁxed scanning order, and
{1,π1,...,π N}are rationally independent, the empirical distribution P(τ)
T converges to the tar-
get distribution πas T →∞ for any τ ≥0.
A set of nreal numbers, x1,x2,...,x n, is said to be rationally independent if for any set of rational
numbers, a1,a2,...,a n, we have ∑n
i=1 aixi = 0⇔ai = 0,∀1 ≤i≤n. The proof of Theorem 1
consists of ﬁrst formulating the dynamics of the weight vector as a constant translation mapping in
a circular unit cube, and then proving that the weights are uniformly distributed by making use of
Kronecker-Weyl’s theorem [25].
For fully-connected (complete) graphs, convergence is guaranteed even with rational ratios. In fact,
herded Gibbs converges to the target joint distribution at a rate ofO(1/T) with a O(log(T)) burn-in
period. This statement is formalized in Theorem 2.
Theorem 2. For a fully-connected graph, when herded Gibbs has a ﬁxed scanning order and a
Dobrushin coefﬁcient of the corresponding Gibbs sampler η <1, there exist constants l >0, and
B >0 such that
dv(P(τ)
T −π) ≤λ
T,∀T ≥T∗,τ >τ∗(T) (7)
where λ= 2N(1+η)
l(1−η) , T∗= 2B
l , τ∗(T) = log2
1+η
(
(1−η)lT
4N
)
, and dv(δπ) :=1
2 ||δπ||1.
The constants land Bare deﬁned in Equation 31 for Proposition 4 in the appendix. If we ignore the
burn-in period and start collecting samples simply from the beginning, we achieve a convergence rate
of O(log(T)
T ) as stated in Corollary 10 in the appendix. The constantlin the convergence rate has an
exponential term, with N in the exponent. An exponentially large constant seems to be unavoidable
for any sampling algorithm when considering the convergence to a joint distribution with2N states.
As for the marginal distributions, it is obvious that the convergence rate of herded Gibbs is also
4
O(1/T) because marginal probabilities are linear functions of the joint distribution. However, in
practice, we observe very rapid convergence results for the marginals, so stronger theoretical results
about the convergence of the marginal distributions seem plausible.
The proof proceeds by ﬁrst bounding the discrepancy between the chain of empirical estimates of
the joint obtained by averaging over T herded Gibbs samples, {P(s)
T },s ≥τ, and a Gibbs chain
initialized at P(τ)
T . After one iteration, this discrepancy is bounded above by O(1/T).
The Gibbs chain has geometric convergence to π and the distance between the Gibbs and herded
Gibbs chains is bounded by O(1/T). The geometric convergence rate to πdominates the discrep-
ancy of herded Gibbs and thus we infer that P(τ)
T converges to π geometrically. To round-off the
proof, we must ﬁnd a limiting value for τ. The proof concludes with an O(log(T)) burn-in for τ.
However, for a generic graph we have no mathematical guarantees on the convergence rate of herded
Gibbs. In fact, one can easily construct synthetic examples for which herded Gibbs does not seem
to converge to the true marginals and joint distribution. For the examples covered by our theorems
and for examples with real data, herded Gibbs demonstrates good behaviour. The exact conditions
under which herded Gibbs converges for sparsely connected graphs are still unknown.
4 Experiments
4.1 Simple Complete Graph
We begin with an illustration of how herded Gibbs substantially outperforms Gibbs on a simple
complete graph. In particular, we consider a fully-connected model of two variables, X1 and X2,
as shown in Figure 2; the joint distribution of these variables is shown in Table 1. Figure 3 shows
the marginal distribution P(X1 = 1) approximated by both Gibbs and herded Gibbs for different
ϵ. As ϵ decreases, both approaches require more iterations to converge, but herded Gibbs clearly
outperforms Gibbs. The ﬁgure also shows that Herding does indeed exhibit a linear convergence
rate.
X1 X2
Figure 2: Two-variable model.
X1 = 0 X 1 = 1 P(X2)
X2 = 0 1/4 −ϵ ϵ 1/4
X2 = 1 ϵ 3/4 −ϵ 3/4
P(X1) 1/4 3 /4 1
Table 1: Joint distribution of the two-variable model.
4.2 MRF for Image Denoising
Next, we consider the standard setting of a grid-lattice MRF for image denoising. Let us assume
that we have a binary image corrupted by noise, and that we want to infer the original clean image.
Let Xi ∈{−1,+1}denote the unknown true value of pixel i, and yi the observed, noise-corrupted
value of this pixel. We take advantage of the fact that neighboring pixels are likely to have the same
label by deﬁning an MRF with an Ising prior. That is, we specify a rectangular 2D lattice with the
following pair-wise clique potentials:
ψij(xi,xj) =
(
eJij e−Jij
e−Jij eJij
)
(8)
and joint distribution:
p(x|J) = 1
Z(J)
∏
i∼j
ψij(xi,xj) = 1
Z(J) exp

1
2
∑
i∼j
Jijxixj

, (9)
where i∼jis used to indicate that nodesiand jare connected. The known parametersJij establish
the coupling strength between nodes iand j. Note that the matrix J is symmetric. If all the Jij >0,
then neighboring pixels are likely to be in the same state.
5
(a) Approximate marginals obtained via Gibbs (blue) and herded Gibbs (red).
(b) Log-log plot of marginal approximation errors obtained via Gibbs (blue) and herded Gibbs (red).
(c) Inverse of marginal approximation errors obtained via Gibbs (blue) and herded Gibbs (red).
Figure 3: (a) Approximating a marginal distribution with Gibbs (blue) and herded Gibbs (red) for an
MRF of two variables, constructed so as to make the move from state (0,0) to (1,1) progressively
more difﬁcult as ϵ decreases. The four columns, from left to right, are for ϵ = 0.1, ϵ = 0.01,
ϵ = 0.001 and ϵ = 0.0001. Table 1 provides the joint distribution for these variables. The error
bars for Gibbs correspond to one standard deviation. Rows (b) and (c) illustrate that the empirical
convergence rate of herded Gibbs matches the expected theoretical rate. In the plots of rows (b)
and (c), the upper-bound in the error of herded Gibbs was used to remove the oscillations so as to
illustrate the behaviour of the algorithm more clearly.
The MRF model combines the Ising prior with a likelihood model as follows:
p(x,y) = p(x)p(y|x) =

1
Z
∏
i∼j
ψij(xi,xj)

.
[∏
i
p(yi|xi)
]
(10)
The potentials ψij encourage label smoothness. The likelihood terms p(yi|xi) are conditionally
independent (e.g. Gaussians with known variance σ2 and mean µ centered at each value of xi,
denoted µxi). In more precise terms,
p(x,y|J,µ,σ) = 1
Z(J,µ,σ) exp

1
2
∑
i∼j
Jijxixj − 1
2σ2
∑
i
(yi −µxi)2

. (11)
6
Figure 4: Original image (left) and its corrupted version (right), with noise parameter σ= 4.
Figure 5: Reconstruction errors for the image denoising task. The results are averaged across
10 corrupted images with Gaussian noise N(0,16). The error bars correspond to one standard
deviation. Mean ﬁeld requires the speciﬁcation of the damping factor D.
When the coupling parameters Jij are identical, say Jij = J, we have ∑
ijJijf(xi,xj) =
J∑
ijf(xi,xj). Hence, different neighbor conﬁgurations result in the same value of
J∑
ijf(xi,xj). If we store the conditionals for conﬁgurations with the same sum together, we
only need to store as many conditionals as different possible values that the sum could take. This
enables us to develop a shared version of herded Gibbs that is more memory efﬁcient where we only
maintain and update weights for distinct states of the Markov blanket of each variable.
In this exemplary image denoising experiment, noisy versions of the binary image, seen in Figure 4
(left), were created through the addition of Gaussian noise, with varying σ. Figure 4 (right) shows a
corrupted image with σ= 4. The L2 reconstruction errors as a function of the number of iterations,
for this example, are shown in Figure 5. The plot compares the herded Gibbs method against Gibbs
and two versions of mean ﬁeld with different damping factors [19]. The results demonstrate that the
herded Gibbs techiques are among the best methods for solving this task.
A comparison for different valuesσis presented in Table 2. As expected mean ﬁeld does well in the
low-noise scenario, but the performance of the shared version of herded Gibbs as the noise increases
is signiﬁcantly better.
7
Table 2: Errors of image denoising example after 30 iterations (all measurements have been scaled
by ×10−3). We use an Ising prior with Jij = 1 and four Gaussian noise models with different
σ’s. For each σ, we generated 10 corrupted images by adding Gaussian noise. The ﬁnal results
shown here are averages and standard deviations (in parentheses) across the 10 corrupted images. D
denotes the damping factor in mean ﬁeld.
PPPPPPMethod
σ 2 4 6 8
Herded Gibbs 21.58(0.26) 32 .07(0.98) 47 .52(1.64) 67 .93(2.78)
Herded Gibbs - shared 22.24(0.29) 31.40(0.59) 42.62(1.98) 58.49(2.86)
Gibbs 21.63(0.28) 37 .20(1.23) 63 .78(2.41) 90 .27(3.48)
Mean ﬁeld (D=0.5) 15.52(0.30) 41 .76(0.71) 76 .24(1.65) 104 .08(1.93)
Mean ﬁeld (D=1) 17.67(0.40) 32 .04(0.76) 51 .19(1.44) 74 .74(2.21)
Other
’
Person
Zed
Other
s
Other
.
Other
’
Person
Zed
Other
s
Other
dead
Other
.
Other
dead
Other
baby
Figure 6: Typical skip-chain CRF model for named entity recognition.
4.3 CRF for Named Entity Recognition
Named Entity Recognition (NER) involves the identiﬁcation of entities, such as people and loca-
tions, within a text sample. A conditional random ﬁed (CRF) for NER models the relationship
between entity labels and sentences with a conditional probability distribution: P(Y|X,θ), where
Xis a sentence, Y is a labeling, andθis a vector of coupling parameters. The parameters,θ, are fea-
ture weights and model relationships between variables Yi and Xj or Yi and Yj. A chain CRF only
employs relationships between adjacent variables, whereas a skip-chain CRF can employ relation-
ships between variables where subscriptsiand jdiffer dramatically. Skip-chain CRFs are important
in language tasks, such as NER and semantic role labeling, because they allow us to model long
dependencies in a stream of words, see Figure 6.
Once the parameters have been learned, the CRF can be used for inference; a labeling for some
sentence X is found by maximizing the above probability. Inference for CRF models in the NER
domain is typically carried out with the Viterbi algorithm. However, if we want to accommodate long
term dependencies, thus resulting in the so called skip-chain CRFs, Viterbi becomes prohibitively
expensive. To surmount this problem, the Stanford named entity recognizer [15] makes use of
annealed Gibbs sampling.
To demonstrate herded Gibbs on a practical application of great interest in text mining, we modify
the standard inference procedure of the Stanford named entity recognizer by replacing the annealed
Gibbs sampler with the herded Gibbs sampler. The herded Gibbs sampler in not annealed. To ﬁnd
the maximum a posteriori sequence Y, we simply choose the sample with highest joint discrete
probability. In order to be able to compare against Viterbi, we have purposely chosen to use single-
chain CRFs. We remind the reader, however, that the herded Gibbs algorithm could be used in cases
where Viterbi inference is not possible.
We used the pre-trained 3-class CRF model in the Stanford NER package [15]. This model is a
linear chain CRF with pre-deﬁned features and pre-trained feature weights, θ. For the test set, we
used the corpus for the NIST 1999 IE-ER Evaluation. Performance is measured in per-entity F1(
F1 = 2· precision·recall
precision+recall
)
. For all the methods, except Viterbi, we show F1 scores after 100, 400 and
800 iterations in Table 3. For Gibbs, the results shown are the averages and standard deviations
over 5 random runs. We used a linear annealing schedule for Gibbs. As the results illustrate,
8
”Pumpkin” (Tim Roth) and ”Honey Bunny” (Amanda Plummer) are having breakfast in a
diner. They decide to rob it after realizing they could make money off the customers as
well as the business, as they did during their previous heist. Moments after they initiate
the hold-up, the scene breaks off and the title credits roll. As Jules Winnﬁeld (Samuel L.
Jackson) drives, Vincent Vega(John Travolta) talks about his experiences in Europe, from
where he has just returned: the hash bars in Amsterdam, the French McDonald’s and its
”Royale with Cheese”.
Figure 7: Results for the application of the NER CRF to a random Wikipedia sample [1]. Entities
are automatically classiﬁed as Person, Location and Organization.
herded Gibbs attains the same accuracy as Viterbi and it is faster than annealed Gibbs. Unlike
Viterbi, herded Gibbs can be easily applied to skip-chain CRFs. After only 400 iterations (90.5
seconds), herded Gibbs already achieves anF1 score of 84.75, while Gibbs, even after 800 iterations
(115.9 seconds) only achieves an F1 score of 84.61. The experiment thus clearly demonstrates that
(i) herded Gibbs does no worse than the optimal solution, Viterbi, and (ii) herded Gibbs yields
more accurate results for the same amount of computation. Figure 7 provides a representative NER
example of the performance of Gibbs, herded Gibbs and Viterbi (all methods produced the same
annotation for this short example).
Table 3: Gibbs, herded Gibbs and Viterbi for the NER task. The average computational time each
approach took to do inference for the entire test set is listed (in square brackets). After only 400
iterations (90.48 seconds), herded Gibbs already achieves an F1 score of 84.75, while Gibbs, even
after 800 iterations (115.92 seconds) only achieves anF1 score of 84.61. For the same computation,
herded Gibbs is more accurate than Gibbs.
``````````Method
Iterations 100 400 800
Annealed Gibbs 84.36(0.16) [55.73s] 84.51(0.10) [83.49s] 84.61(0.05) [115.92s]
Herded Gibbs 84.70 [59.08s] 84.75 [90.48s] 84.81 [132.00s]
Viterbi 84.81[46.74s]
5 Conclusions and Future Work
In this paper, we introduced herded Gibbs, a deterministic variant of the popular Gibbs sampling al-
gorithm. While Gibbs relies on drawing samples from the full-conditionals at random, herded Gibbs
generates the samples by matching the full-conditionals. Importantly, the herded Gibbs algorithm is
very close to the Gibbs algorithm and hence retains its simplicity of implementation.
The synthetic, denoising and named entity recognition experiments provided evidence that herded
Gibbs outperforms Gibbs sampling. However, as discussed, herded Gibbs requires storage of the
conditional distributions for all instantiations of the neighbors in the worst case. This storage re-
quirement indicates that it is more suitable for sparse probabilistic graphical models, such as the
CRFs used in information extraction. At the other extreme, the paper advanced the theory of de-
terministic sampling by showing that herded Gibbs converges with rate O(1/T) for models with
independent variables and fully-connected models. Thus, there is gap between theory and practice
that needs to be narrowed. We do not anticipate that this will be an easy task, but it is certainly a key
direction for future work.
We should mention that it is also possible to design parallel versions of herded Gibbs in a Jacobi
fashion. We have indeed studied this and found that these are less efﬁcient than the Gauss-Seidel
version of herded Gibbs discussed in this paper. However, if many cores are available, we strongly
recommend the Jacobi (asynchronous) implementation as it will likely outperform the Gauss-Seidel
(synchronous) implementation.
9
The design of efﬁcient herding algorithms for densely connected probabilistic graphical models
remains an important area for future research. Such algorithms, in conjunction with Rao Black-
wellization, would enable us to attack many statistical inference tasks, including Bayesian variable
selection and Dirichlet processes.
There are also interesting connections with other algorithms to explore. If, for a fully connected
graphical model, we build a new graph where every state is a node and directed connections exist
between nodes that can be reached with a single herded Gibbs update, then herded Gibbs becomes
equivalent to the Rotor-Router model of Alex Holroyd and Jim Propp1 [13]. This deterministic ana-
logue of a random walk has provably superior concentration rates for quantities such as normalized
hitting frequencies, hitting times and occupation frequencies. In line with our own convergence
results, it is shown that discrepancies in these quantities decrease as O(1/T) instead of the usual
O(1/
√
T). We expect that many of the results from this literature apply to herded Gibbs as well.
The connection with the work of Art Owen and colleagues, see for example [7], also needs to
be explored further. Their work uses completely uniformly distributed (CUD) sequences to drive
Markov chain Monte Carlo schemes. It is not clear, following discussions with Art Owen, that CUD
sequences can be constructed in a greedy way as in herding.
References
[1] Pulp ﬁction - wikipedia, the free encyclopedia @ONLINE, June 2012.
[2] D. H. Ackley, G. Hinton, and T.. Sejnowski. A learning algorithm for Boltzmann machines.
Cognitive Science, 9:147–169, 1985.
[3] C. Andrieu, N. de Freitas, A. Doucet, and M. I. Jordan. An Introduction to MCMC for Machine
Learning. Machine Learning, 50(1):5–43, 2003.
[4] F. Bach, S. Lacoste-Julien, and G. Obozinski. On the equivalence between herding and condi-
tional gradient algorithms. In International Conference on Machine Learning, 2012.
[5] P. Br ´emaud. Markov chains: Gibbs ﬁelds, Monte Carlo simulation, and queues , volume 31.
Springer, 1999.
[6] P. Carbonetto, J. Kisynski, N. de Freitas, and D. Poole. Nonparametric Bayesian logic. In
Uncertainty in Artiﬁcial Intelligence, pages 85–93, 2005.
[7] S. Chen, J. Dick, and A. B. Owen. Consistency of Markov chain quasi-Monte Carlo on con-
tinuous state spaces. Annals of Statistics, 39(2):673–701, 2011.
[8] Y . Chen, M. Welling, and A.J. Smola. Supersamples from kernel-herding. In Uncertainty in
Artiﬁcial Intelligence, pages 109–116, 2010.
[9] A. Doucet, N. de Freitas, and N. Gordon. Sequential Monte Carlo Methods in Practice. Statis-
tics for Engineering and Information Science. Springer, 2001.
[10] A. Gelfand, Y . Chen, L. van der Maaten, and M. Welling. On herding and the perceptron
cycling theorem. In Advances in Neural Information Processing Systems , pages 694–702,
2010.
[11] N. D. Goodman, V . K. Mansinghka, D. M. Roy, K. Bonawitz, and J. B. Tenenbaum. Church:
a language for generative models. Uncertainty in Artiﬁcial Intelligence, 2008.
[12] G.E. Hinton and R.R. Salakhutdinov. Reducing the dimensionality of data with neural net-
works. Science, 313(5786):504–507, 2006.
[13] Alexander E Holroyd and James Propp. Rotor walks and Markov chains. Algorithmic Proba-
bility and Combinatorics, 520:105–126, 2010.
[14] F. Husz ´ar and D. Duvenaud. Optimally-weighted herding is Bayesian quadrature. Arxiv
preprint arXiv:1204.1664, 2012.
[15] T. Grenager J. R. Finkel and C. Manning. Incorporating non-local information into informa-
tion extraction systems by Gibbs sampling. In Proceedings of the 43rd Annual Meeting on
Association for Computational Linguistics, ACL ’05, pages 363–370, Stroudsburg, PA, USA,
2005. Association for Computational Linguistics.
1We thank Art Owen for pointing out this connection.
10
[16] J. S. Liu. Monte Carlo strategies in scientiﬁc computing. Springer, 2001.
[17] D. J. Lunn, A. Thomas, N. Best, and D. Spiegelhalter. WinBUGS a Bayesian modelling frame-
work: Concepts, structure, and extensibility. Statistics and Computing, 10(4):325–337, 2000.
[18] B. Milch and S. Russell. General-purpose MCMC inference over relational structures. In
Uncertainty in Artiﬁcial Intelligence, pages 349–358, 2006.
[19] K. P. Murphy. Machine Learning: a Probabilistic Perspective. MIT Press, 2012.
[20] I. Murray and L. T. Elliott. Driving Markov chain Monte Carlo with a dependent random
stream. Technical Report arXiv:1204.3187, 2012.
[21] I. Porteous, D. Newman, A. Ihler, A. Asuncion, P. Smyth, and M. Welling. Fast collapsed
Gibbs sampling for latent Dirichlet allocation. In ACM SIGKDD international conference on
Knowledge discovery and data mining, pages 569–577, 2008.
[22] C. P. Robert and G. Casella. Monte Carlo Statistical Methods. Springer, 2nd edition, 2004.
[23] M. Welling. Herding dynamic weights for partially observed random ﬁeld models. In Uncer-
tainty in Artiﬁcial Intelligence, pages 599–606, 2009.
[24] M. Welling. Herding dynamical weights to learn. In International Conference on Machine
Learning, pages 1121–1128, 2009.
[25] Hermann Weyl. ¨Uber die gleichverteilung von zahlen mod. eins. Mathematische Annalen,
77:313–352, 1916.
11
A Proof of Theorem 1
We ﬁrst show that the weight dynamics of a one-variable herding algorithm are restricted to an
invariant interval of length 1.
Lemma 3. If wis the weight of the herding dynamics of a single binary variableXwith probability
P(X = 1) =π, and w(s) ∈(π−1,π] at some step s≥0, then w(t) ∈(π−1,π],∀t≥s. Moreover,
for T ∈N, we have:
s+T∑
t=s+1
I[X(t) = 1]∈[Tπ −1,Tπ + 1] (12)
s+T∑
t=s+1
I[X(t) = 0]∈[T(1 −π) −1,T(1 −π) + 1]. (13)
Proof. We ﬁrst show that w ∈(π−1,π],∀t ≥s. This is easy to observe by induction as w(s) ∈
(π−1,π] and if w(t) ∈(π−1,π] for some t≥s, then, following Equation 5, we have:
w(t+1) =
{
w(t) + π−1 ∈(π−1,2π−1] ⊆(π−1,π] if w(t) >0
w(t) + π∈(2π−1,π] ⊆(π−1,π] otherwise. (14)
Summing up both sides of Equation 5 over timmediately gives us the result of Equation 12 since:
Tπ −
s+T∑
t=s+1
I[X(t) = 1] =w(s+T) −w(s) ∈[−1,1]. (15)
In addition, Equation 13 follows by observing that I[X(t) = 0] = 1−I[X(t) = 1].
When wis outside the invariant interval, it is easy to observe thatwwill move into it monotonically
at a linear speed in a transient period. So we will always consider an initialization ofw∈(π−1,π]
from now on.
Equivalently, we can take a one-to-one mapping w ←w mod 1(we deﬁne 1 mod 1 = 1) and
think of w as updated by a constant translation vector in a circular unit interval (0,1] as shown in
Figure 8. That is,
w(t) = (w(t−1) + π) mod 1 , x (t) =
{
1 if w(t−1) <π
0 otherwise (16)
Figure 8: Equivalent weight dynamics for
a single variable.
Figure 9: Dynamics of herding with two indepen-
dent variables.
We are now ready to give the proof of Theorem 1.
12
Proof of Theorem 1. For an empty graph of N independent vertices, the dynamics of the weight
vector w are equivalent to a constant translation mapping in an N-dimensional circular unit space
(0,1], as shown in Figure 9:
w(t) = ( w(t−1) + π) mod 1
= ( w(0) + tπ) mod 1 , x (t)
i =
{
1 if w(t−1)
i <πi
0 otherwise ,∀1 ≤i≤N (17)
The Kronecker-Weyl theorem [25] states that the sequence ˜w(t) = tπ mod 1,t ∈Z+ is equidis-
tributed (or uniformly distributed) on (0,1] if and only if (1,π1,...,π N) is rationally indepen-
dent. Since we can deﬁne a one-to-one volume preserving transformation between ˜w(t) and w(t)
as ( ˜w(t) + w(0)) mod 1 =w(t), the sequence of weights {w(t)}is also uniformly distributed in
(0,1]N.
Deﬁne the mapping from a state value xi to an interval of wi as
Ai(x) =
{
(0,πi] if x= 1
(πi,1] if x= 0 (18)
and let |Ai|be its measure. We obtain the limiting distribution of the joint state as
lim
T→∞
P(τ)
T (X = x) = lim
T→∞
1
T
T∑
t=1
I
[
w(t−1) ∈
N∏
i=1
Ai(xi)
]
=
N∏
i=1
|Ai(xi)|
=
N∏
i=1
π(Xi = xi)
= π(X = x) (19)
13
B Proof of Theorem 2
In this appendix, we give an upper bound for the convergence rate of the sampling distribution in
fully connected graphs. As herded Gibbs sampling is deterministic, the distribution of a variable’s
state at every iteration degenerates to a single state. As such, we study here the empirical distribution
of a collection of samples.
The structure of the proof is as follows (with notation deﬁned in the next subsection): We study the
distribution distance between the invariant distributionπand the empirical distribution ofT samples
collected starting from sweep τ, P(τ)
T . We show that the distance decreases as τ ⇒τ + 1with the
help of an auxiliary regular Gibbs sampling Markov chain initialized at π(0) = P(τ)
T , as shown in
Figure 10. On the one hand, the distance between the regular Gibbs chain after one iteration, π(1),
and πdecreases according to the geometric convergence property of MCMC algorithms on compact
state spaces. On the other hand, we show that in one step the distance between P(τ+1)
T and π(1)
increases by at most O(1/T). Since the O(1/T) distance term dominates the exponentially small
distance term, the distance between P(τ+1)
T and π is bounded by O(1/T). Moreover, after a short
burn-in period, L= O(log(T)), the empirical distribution P(τ+L)
T will have an approximation error
in the order of O(1/T).
Figure 10: Transition kernels and relevant distances for the proof of Theorem 2.
B.1 Notation
Assume without loss of generality that in the systematic scanning policy, the variables are sampled
in the order 1,2,··· ,N.
B.1.1 State Distribution
• Denote by X+ the support of the distribution π, that is, the set of states with positive
probability.
• We use τ to denote the time in terms of sweeps over all of the N variables, and tto denote
the time in terms of steps where one step constitutes the updating of one variable. For
example, at the end of τ sweeps, we have t= τN.
• Recall the sample/empirical distribution, P(τ)
T , presented in Deﬁnition 1. Figure 11 pro-
vides a visual interpretation of the deﬁnition.
•Denote the sample/empirical distribution at theithstep within a sweep asP(τ)
T,i ,τ ≥0,T >
0,0 ≤i≤N, as shown in Figure 12:
P(τ)
T,i (X = x) = 1
T
τ+T−1∑
k=τ
I(X(kN+i) = x).
14
Figure 11: Distribution over time at the end of every sweep.
This is the distribution of T samples collected at the ith step of every sweep, starting from
the τth sweep. Clearly, P(τ)
T = P(τ)
T,0 = P(τ−1)
T,N .
Figure 12: Distribution over time within a sweep.
•Denote the distribution of a regular Gibbs sampling Markov chain afterLsweeps of updates
over the N variables with π(L),L ≥0.
For a given time τ, we construct a Gibbs Markov chain with initial distribution π0 = P(τ)
T
and the same scanning order of herded Gibbs, as shown in Figure 10.
B.1.2 Transition Kernel
•Denote the transition kernel of regular Gibbs for the step of updating a variableXi with Ti,
and for a whole sweep with T.
By deﬁnition, π0T = π1. The transition kernel for a single step can be represented as a
2N ×2N matrix:
Ti(x,y) =
{
0 if x−i ̸= y−i
π(Xi = yi|x−i) otherwise ,1 ≤i≤N,x,y ∈{0,1}N (20)
where x is the current state vector of N variables, y is the state of the next step, and x−i
denotes all the components of x excluding the ith component. If π(x−i) = 0, the condi-
tional probability is undeﬁned and we set it with an arbitrary distribution. Consequently,T
can also be represented as:
T = T1T2 ···T N.
•Denote the Dobrushin ergodic coefﬁcient [5] of the regular Gibbs kernel with η ∈[0,1].
When η <1, the regular Gibbs sampler has a geometric rate of convergence of
dv(π(1) −π) =dv(Tπ(0) −π) ≤ηdv(π(0) −π),∀π(0). (21)
A common sufﬁcient condition for η <1 is that π(X) is strictly positive.
•Consider the sequence of sample distributionsP(τ)
T ,τ = 0,1,··· in Figures 11 and 12. We
deﬁne the transition kernel of herded Gibbs for the step of updating variable Xi with ˜T(τ)
T,i ,
and for a whole sweep with ˜T(τ)
T .
Unlike regular Gibbs, the transition kernel is not homogeneous. It depends on both the
time τ and the sample size T. Nevertheless, we can still represent the single step transition
15
kernel as a matrix:
˜T(τ)
T,i (x,y) =
{ 0 if x−i ̸= y−i
P(τ)
T,i (Xi = yi|x−i) if x−i = y−i
,1 ≤i≤N,x,y ∈{0,1}N,
(22)
where P(τ)
T,i (Xi = yi|x−i) is deﬁned as:
P(τ)
T,i (Xi = yi|x−i) =Nnum
Nden
Nnum = TP(τ)
T,i (X−i = x−i,Xi = yi) =
τ+T−1∑
k=τ
I(X(kN+i)
−i = x−i,X(kN+i)
i = yi)
Nden = TP(τ)
T,i−1(X−i = x−i) =
τ+T−1∑
k=τ
I(X(kN+i−1)
−i = x−i), (23)
where Nnum is the number of occurrences of a joint state, and Nden is the number of oc-
currences of a conditioning state in the previous step. When π(x−i) = 0, we know that
Nden = 0 with a proper initialization of herded Gibbs, and we simply set ˜T(τ)
T,i = Ti for
these entries. It is not hard to verify the following identity by expanding every term with
its deﬁnition
P(τ)
T,i = P(τ)
T,i−1 ˜T(τ)
T,i
and consequently,
P(τ+1)
T = P(τ)
T ˜T(τ)
T
with
˜T(τ)
T = ˜T(τ)
T,1 ˜T(τ)
T,2 ··· ˜T(τ)
T,N.
B.2 Linear Visiting Rate
We prove in this section that every joint state in the support of the target distribution is visited, at
least, at a linear rate. This result will be used to measure the distance between the Gibbs and herded
Gibbs transition kernels.
Proposition 4. If a graph is fully connected, herded Gibbs sampling scans variables in a ﬁxed order,
and the corresponding Gibbs sampling Markov chain is irreducible, then for any state x ∈X+ and
any index i∈[1,N], the state is visited at least at a linear rate. Speciﬁcally,
∃l> 0,B >0,s.t., ∀i∈[1,N],x ∈X+,T ∈N,s ∈N
s+T−1∑
k=s
I
[
X(t=Nk+i) = x
]
≥lT −B (24)
Denote the minimum nonzero conditional probability as
πmin = min
1≤i≤N,π(xi|x−i)>0
π(xi|x−i).
The following lemma, which is needed to prove Proposition 4, gives an inequality between the
number of visits of two sets of states in consecutive steps.
Lemma 5. For any integeri∈[1,N] and two sets of states X,Y ⊆X+ with a mapping F : X →Y
that satisﬁes the following condition:
∀x ∈X,F(x)−i = x−i, ∪x∈XF(x) =Y, (25)
we have that, for any s ≥ 0 and T > 0, the number of times Y is visited in the set of steps
Ci = {t= kN + i: s≤k≤k+ T −1}is lower bounded by a function of the number of times X
is visited in the previous steps Ci−1 = {t= kN + i−1 :s≤k≤k+ T −1}as:
∑
t∈Ci
I
[
X(t) ∈Y
]
≥πmin
∑
t∈Ci−1
I
[
X(t) ∈X
]
−|Y| (26)
16
Proof. As a complement to Condition 25, we can deﬁne F−1 as the inverse mapping from Y to
subsets of X so that for any y ∈Y, x ∈F−1(y), we have x−i = y−i, and ∪y∈YF−1(y) =X.
Consider any state y ∈ Y, when y is visited in Ci, the weight wi,y−i is active. Let us denote
the set of all the steps in [sN + 1,s(N + T) +N] when wi,y−i is active by Ci(y−i), that is,
Ci(y−i) ={t: t∈Ci,X(t)
−i = y−i}. Applying Lemma 3 we get
∑
t∈Ci
I
[
X(t) = y
]
≥π(yi|y−i)|Ci(y−i)|−1 ≥πmin|Ci(y−i)|−1. (27)
Since the variables X−i are not changed at steps in Ci, we have
|Ci(y−i)|=
∑
t∈Ci−1
I
[
X(t)
−i = y−i
]
≥
∑
t∈Ci−1
I
[
X(t) ∈F−1(y)
]
. (28)
Combining the fact that ∪y∈YF−1(y) = X and summing up both sides of Equation 27 over Y
proves the lemma:
∑
t∈Ci
I
[
X(t) ∈Y
]
≥
∑
y∈Y

πmin
∑
t∈Ci−1
I
[
X(t) ∈F−1(y)
]
−1

≥πmin
∑
t∈Ci−1
I
[
X(t) ∈X
]
−|Y|.
(29)
Remark 6. A fully connected graph is a necessary condition for the application of Lemma 3 in the
proof. If a graph is not fully connected ( N(i) ̸= −i), a weight wi,yN(i) may be shared by multiple
full conditioning states. In this case Ci(y−i) is no longer a consecutive sequence of times when the
weight is updated, and Lemma 3 does not apply here.
Now let us prove Proposition 4 by iteratively applying Lemma 5.
Proof of Proposition 4. Because the corresponding Gibbs sampler is irreducible and any Gibbs sam-
pler is aperiodic, there exists a constant t∗ > 0 such that for any state y ∈X+, and any step in a
sweep, i, we can ﬁnd a path of length t∗for any state x ∈X+ with a positive transition probability,
Path(x) = (x = x(0),x(1),..., x(t∗) =y), to connect from x to y, where each step of the path
follows the Gibbs updating scheme. For a strictly positive distribution, the minimum value of t∗is
N.
Denote τ∗= ⌈t∗/N⌉and the jth element of the path Path(x) as Path(x,j). We can deﬁne t∗+ 1
subsets Sj ⊆X+,0 ≤j ≤t∗as the union of all the jth states in the path from any state in X+:
Sj = ∪x∈X+ Path(x,j)
By deﬁnition of these paths, we knowS0 = X+ and St∗ = {y}, and there exits an integeri(j) and a
mapping Fj : Sj−1 →Sj,∀jthat satisfy the condition in Lemma 5 (i(j) is the index of the variable
to be updated, and the mapping is deﬁned by the transition path). Also notice that any state in Sj
can be different from y by at most min{N,t∗−j}variables, and therefore |Sj|≤ 2min{N,t∗−j}.
17
Let us apply Lemma 5 recursively from j = t∗to 1 as
s+T−1∑
k=s
I
[
X(t=Nk+i) = y
]
≥
s+T−1∑
k=s+τ∗
I
[
X(t=Nk+i) = y
]
=
s+T−1∑
k=s+τ∗
I
[
X(t=Nk+i) ∈St∗
]
≥πmin
s+T−1∑
k=s+τ∗
I
[
X(t=Nk+i−1) ∈St∗−1
]
−|St∗|
≥···
≥πt∗
min
s+T−1∑
k=s+τ∗
I
[
X(t=Nk+i−t∗) ∈S0 = X+
]
−
t∗−1∑
j=0
πj
min|St∗−j|
≥πt∗
min(T −τ∗) −
t∗−1∑
j=0
πj
min2min{N,j}. (30)
The proof is concluded by choosing the constants
l= πt∗
min, B = τ∗πt∗
min +
t∗−1∑
j=0
πj
min2min{N,j}. (31)
B.3 Herded Gibbs’s Transition Kernel ˜T(τ)
T is an Approximation to T
The following proposition shows that ˜T(τ)
T is an approximation to the regular Gibbs sampler’s tran-
sition kernel T with an error of O(1/T).
Proposition 7. For a fully connected graph, if the herded Gibbs has a ﬁxed scanning order and the
corresponding Gibbs sampling Markov chain is irreducible, then for any τ ≥0, T ≥T∗ := 2B
l
where land Bare the constants in Proposition 4, the following inequality holds:
∥˜T(τ)
T −T∥∞≤4N
lT (32)
Proof. When x ̸∈X+, we have the equality ˜T(τ)
T,i (x,y) = Ti(x,y) by deﬁnition. When x ∈X+
but y ̸∈X+, then Nden = 0 (see the notation of ˜T(τ)
T for deﬁnition of Nden) as y will never be
visited and thus ˜T(τ)
T,i (x,y) = 0 =Ti(x,y) also holds. Let us consider the entries in ˜T(τ)
T,i (x,y)
with x,y ∈X+ in the following.
Because X−i is not updated at ith step of every sweep, we can replacei−1 in the deﬁnition of Nden
by iand get
Nden =
τ+T−1∑
k=τ
I(X(kN+i)
−i = x−i).
Notice that the set of times{t= kN+i: τ ≤k≤τ+T−1,Xt
−i = x−i)}, whose size is Nden, is a
consecutive set of times whenwi,x−i is updated. By Lemma 3, we obtain a bound for the numerator
Nnum ∈[Ndenπ(Xi = yi|x−i) −1,Ndenπ(Xi = yi|x−i) + 1]⇔
|P(τ)
T,i (Xi = yi|x−i) −π(Xi = yi|x−i)|= |Nnum
Nden
−π(Xi = yi|x−i)|≤ 1
Nden
. (33)
Also by Proposition 4, we know every state in X+ is visited at a linear rate, there hence exist
constants l> 0 and B >0, such that the number of occurrence of any conditioning state x−i, Nden,
18
is bounded by
Nden ≥
τ+T−1∑
k=τ
I(X(kN+i) = x) ≥lT −B ≥l
2T, ∀T ≥2B
l . (34)
Combining equations (33) and (34), we obtain
|P(τ)
T,i (Xi = yi|x−i) −π(Xi = yi|x−i)|≤ 2
lT, ∀T ≥2B
l . (35)
Since the matrix ˜T(τ)
T,i and Ti differ only at those elements where x−i = y−i, we can bound the L1
induced norm of the transposed matrix of their difference by
∥( ˜T(τ)
T,i −Ti)T∥1 = max
x
∑
y
|˜T(τ)
T,i (x,y) −Ti(x,y)|
= max
x
∑
yi
|P(τ)
T,i (Xi = yi|x−i) −π(Xi = yi|x−i)|
≤ 4
lT, ∀T ≥2B
l (36)
Observing that both ˜T(τ)
T and T are multiplications of N component transition matrices, and the
transition matrices, ˜T(τ)
T and Ti, have a unit L1 induced norm as:
∥( ˜T(τ)
T,i )T∥1 = max
x
∑
y
|˜T(τ)
T,i (x,y)|= max
x
∑
y
P(τ)
T,i (Xi = yi|x−i) = 1 (37)
∥(Ti)T∥1 = max
x
∑
y
|Ti(x,y)|= max
x
∑
y
P(Xi = yi|x−i) = 1 (38)
we can further bound the L1 norm of the difference, ( ˜T(τ)
T −T)T. Let P ∈RN be any vector with
nonzero norm. Using the triangular inequality, the difference of the resulting vectors after applying
˜T(τ)
T and T is bounded by
∥P( ˜T(τ)
T −T)∥1 =∥P˜T(τ)
T,1 ... ˜T(τ)
T,N −PT... TN∥1
≤∥P˜T(τ)
T,1 ˜T(τ)
T,2 ... ˜T(τ)
T,N −PT1 ˜T(τ)
T,2 ... ˜T(τ)
T,N∥1+
∥PT1 ˜T(τ)
T,2 ˜T(τ)
T,3 ... ˜T(τ)
T,N −PT1T2 ˜T(τ)
T,3 ... ˜T(τ)
T,N∥1+
...
∥PT1 ... TN−1 ˜T(τ)
T,N −PT1 ... TN−1TN∥1 (39)
where the i’th term is
∥PT1 ... Ti−1( ˜T(τ)
T,i −Ti) ˜T(τ)
T,i+1 ... ˜T(τ)
T,N∥1 ≤∥PT1 ... Ti−1( ˜T(τ)
T,i −Ti)∥1 (Unit L1 norm, Eqn. 37)
≤∥PT1 ... Ti−1∥1
4
lT (Eqn. 36)
≤∥P∥1
4
lT (Unit L1 norm, Eqn. 38)
(40)
Consequently, we get the L1 induced norm of ( ˜T(τ)
T −T)T as
∥( ˜T(τ)
T −T)T∥= max
P
∥P( ˜T(τ)
T −T)∥1
∥P∥1
≤4N
lT , ∀T ≥2B
l , (41)
19
B.4 Proof of Theorem 2
When we initialize the herded Gibbs and regular Gibbs with the same distribution (see Figure 10),
since the transition kernel of herded Gibbs is an approximation to regular Gibbs and the distribution
of regular Gibbs converges to the invariant distribution, we expect that herded Gibbs also approaches
the invariant distribution.
Proof of Theorem 2. Construct an auxiliary regular Gibbs sampling Markov chain initialized with
π(0)(X) = P(τ)
T (X) and the same scanning order as herded Gibbs. As η <1, the Gibbs Markov
chain has uniform geometric convergence rate as shown in Equation (21).
Also, the Gibbs Markov chain must be irreducible due to η <1 and therefore Proposition 7 applies
here. We can bound the distance between the distributions of herded Gibbs after one sweep of all
variables, P(τ+1)
T , and the distribution after one sweep of regular Gibbs sampling, π(1) by
dv(P(τ+1)
T −π(1)) =dv(π(0)( ˜T(τ)
T −T)) =1
2∥π(0)( ˜T(τ)
T −T)∥1
≤2N
lT ∥π(0)∥1 = 2N
lT , ∀T ≥T∗,τ ≥0. (42)
Now we study the change of discrepancy between P(τ)
T and πas a function as τ.
Applying the triangle inequality of dv:
dv(P(τ+1)
T −π) =dv(P(τ+1)
T −π(1) + π(1) −π) ≤dv(P(τ+1)
T −π(1)) +dv(π(1) −π)
≤2N
lT + ηdv(P(τ)
T −π), ∀T ≥T∗,τ ≥0. (43)
The last inequality follows Equations (21) and (42). When the sample distribution is outside a
neighborhood of π, Bϵ1 (π), with ϵ1 = 4N
(1−η)lT, i.e.
dv(P(τ)
T −π) ≥ 4N
(1 −η)lT, (44)
we get a geometric convergence rate toward the invariant distribution by combining the two equa-
tions above:
dv(P(τ+1)
T −π) ≤1 −η
2 dv(P(τ)
T −π) +ηdv(P(τ)
T −π) =1 +η
2 dv(P(τ)
T −π). (45)
So starting from τ = 0, we have a burn-in period for herded Gibbs to enterBϵ1 (π) in a ﬁnite number
of rounds. Denote the ﬁrst time it enters the neighborhood by τ′. According to the geometric
convergence rate in Equations 45 and dv(P(0)
T −π) ≤1
τ′≤
⌈
log1+η
2
( ϵ1
dv(P(0)
T −π)
)
⌉
≤
⌈
log1+η
2
(ϵ1)
⌉
= ⌈τ∗(T)⌉. (46)
After that burn-in period, the herded Gibbs sampler will stay within a smaller neighborhood,Bϵ2 (π),
with ϵ2 = 1+η
1−η
2N
lT , i.e.
dv(P(τ)
T −π) ≤1 +η
1 −η
2N
lT , ∀τ >τ′. (47)
This is proved by induction:
1. Equation (47) holds at τ = τ′+ 1. This is because P(τ′)
T ∈Bϵ1 (π) and following Eqn. 43
we get
dv(P(τ′+1)
T −π) ≤2N
lT + ηϵ1 = ϵ2 (48)
2. For any τ ≥τ′+ 2, assume P(τ−1)
T ∈Bϵ2 (π). Since ϵ2 < ϵ1, P(τ−1)
T is also in the ball
Bϵ1 (π). We can apply the same computation as when τ = τ′+ 1to prove dv(P(τ)
T −π) ≤
ϵ2. So inequality (47) is always satisﬁed by induction.
20
Consequently, Theorem 2 is proved when combining (47) with the inequality τ′ ≤ ⌈τ∗(T)⌉in
Equation( 46).
Remark 8. Similarly to the regular Gibbs sampler, the herded Gibbs sampler also has a burn-in
period with geometric convergence rate. After that, the distribution discrepancy is in the order of
O(1/T), which is faster than the regular Gibbs sampler. Notice that the length of the burn-in period
depends on T, speciﬁcally as a function of log(T).
Remark 9. Irrationality is not required to prove the convergence on a fully-connected graph.
Corollary 10. When the conditions of Theorem 2 hold, and we start collecting samples at the end
of every sweep from the beginning, the error of the sample distribution is bounded by:
dv(P(τ=0)
T −π) ≤λ+ τ∗(T)
T = O(log(T)
T ), ∀T ≥T∗+ τ∗(T∗) (49)
Proof. Since τ∗(T) is a monotonically increasing function of T, for any T ≥T∗+ τ∗(T∗), we can
ﬁnd a number tso that
T = t+ τ∗(t),t ≥T∗.
Partition the sample sequence S0,T = {X(kN) : 0 ≤k < T}into two parts: the burn-in period
S0,τ∗(t) and the stable period Sτ∗(t),T. The discrepancy in the burn-in period is bounded by 1 and
according to Theorem 2, the discrepancy in the stable period is bounded by
dv( ˜P(St,T) −π) ≤λ
t.
Hence, the discrepancy of the whole set S0,T is bounded by
dv( ˜P(S0,T) −π) =dv
(τ∗(t)
T
˜P(S0,τ∗(t)) + t
T
˜P(Sτ∗(t),T) −π
)
≤dv
(τ∗(t)
T ( ˜P(S0,τ∗(t)) −π
)
+ dv
(t
T( ˜P(Sτ∗(t),T) −π
)
≤τ∗(t)
T dv( ˜P(S0,τ∗(t)) −π) + t
Tdv( ˜P(Sτ∗(t),T) −π)
≤τ∗(t)
T ·1 + t
T
λ
t ≤τ∗(T) +λ
T . (50)
21


Peer review:
[Review]:
Nando asked me for some comments and then he thought I should share them on openreview. So here they are.

There are a few other efforts at replacing the IID numbers which drive MCMC. It would be interesting to explore the connections among them.  Here is a sample:

Jim Propp and others have been working on rotor-routers for quite a while. Here is one link:
http://front.math.ucdavis.edu/0904.4507

I've been working with several people on replacing IID numbers by completely uniformly distributed (CUD) ones. This is like taking a small random number generator and using it all up.  See this thesis by Su Chen for the latest results, and lots of references:
www-stat.stanford.edu/~owen/students/SuChenThesis.pdf
or for earlier work, this thesis by Seth Tribble:
www-stat.stanford.edu/~owen/students/SethTribbleThesis.pdf

The oldest papers in that line of work go back to the late 1960s and early 1970s by Chentsov and also by Sobol'.

There is some very recent work by Dick Rudolf and Zhu:
http://arxiv.org/abs/1303.2423
that is similar to herding. The idea there is to make a followup sample of values that fill in holes left after a first sampling.

Not quite as close to this work but still related is the array-RQMC work of Pierre L'Ecuyer and others. See for instance:
www.iro.umontreal.ca/~lecuyer/myftp/papers/mcqmc08-array.pdf