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Exact statistical inference for the Wasserstein distance by selective inference

Selective Inference for the Wasserstein Distance

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Abstract

In this paper, we study statistical inference for the Wasserstein distance, which has attracted much attention and has been applied to various machine learning tasks. Several studies have been proposed in the literature, but almost all of them are based on asymptotic approximation and do not have finite-sample validity. In this study, we propose an exact (non-asymptotic) inference method for the Wasserstein distance inspired by the concept of conditional selective inference (SI). To our knowledge, this is the first method that can provide a valid confidence interval (CI) for the Wasserstein distance with finite-sample coverage guarantee, which can be applied not only to one-dimensional problems but also to multi-dimensional problems. We evaluate the performance of the proposed method on both synthetic and real-world datasets.

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Notes

  1. To make a distinction between random variables and observed variables, we use superscript \(^\mathrm{{obs}}\), e.g., \({\varvec{X}}\) is a random vector and \({\varvec{X}}^\mathrm{{obs}}\) is the observed data vector.

  2. We note that there always exists exactly one redundant equality constraint in linear equality constraint system in (7). This is due to the fact that sum of all the masses on \({\varvec{X}}^\mathrm{{obs}}\) is always equal to sum of all the masses on \({\varvec{Y}}^\mathrm{{obs}}\) (i.e., they are all equal to 1). Therefore, any equality constraint can be expressed as a linear combination of the others, and hence any one constraint can be dropped. In this paper, we always drop the last equality constraint (i.e., the last row of matrix S and the last element of vector \({\varvec{h}}\)) before solving (7).

  3. We suppose that the LP is non-degenerate. A careful discussion might be needed in the presence of degeneracy.

  4. We used dataset Lung_GSE7670 which is available at https://sbcb.inf.ufrgs.br/cumida.

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Acknowledgements

We thank the authors of Imaizumi et al. (2019) for providing us with their software implementation. This work was partially supported by MEXT KAKENHI (20H00601, 16H06538), JST CREST (JPMJCR21D3), JST Moonshot R&D (JPMJMS2033-05), NEDO (JPNP18002, JPNP20006), RIKEN Center for Advanced Intelligence Project, and RIKEN Junior Research Associate Program.

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Authors and Affiliations

  1. Nagoya Institute of Technology, Gokiso-cho, Showa-ku, Nagoya, Aichi, 466-8555, Japan

    Vo Nguyen Le Duy

  2. RIKEN, 2-1 Hirosawa, Wako, Saitama, 351-0198, Japan

    Vo Nguyen Le Duy & Ichiro Takeuchi

  3. Nagoya University, Furo-cho, Chikusa-ku, Nagoya, 464-8601, Japan

    Ichiro Takeuchi

Authors
  1. Vo Nguyen Le Duy
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  2. Ichiro Takeuchi
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Contributions

We provide an exact (non-asymptotic) inference method for the Wasserstein distance. To our knowledge, this is the first method that can provide valid CI, called selective CI, for the Wasserstein distance that guarantees the coverage property in finite sample size. Another practically important advantage of this study is that the proposed method is valid when the Wasserstein distance is computed in multi-dimensional problem, which is impossible for almost all the existing asymptotic methods since the limit distribution of the Wasserstein distance is only applicable for univariate data. We conduct experiments on both synthetic and real-world datasets to evaluate the performance of the proposed method. Our implementation is available at https://github.com/vonguyenleduy/selective_inference_wasserstein_distance.

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Correspondence to Ichiro Takeuchi.

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Appendices

A Proposed method in hypothesis testing framework

We present the proposed method in the setting of hypothesis testing and consider the case when the cost matrix is defined by using squared \(\ell _2\) distance.

Cost matrix We define the cost matrix \(C({\varvec{X}}, {\varvec{Y}})\) of pairwise distances (squared \(\ell _2\) distance) between elements of \({\varvec{X}}\) and \({\varvec{Y}}\) as

$$\begin{aligned} C({\varvec{X}}, {\varvec{Y}})&= \big [(x_i - y_j)^2 \big ]_{ij} \in \mathbb {R}^{n \times m}. \end{aligned}$$
(34)

Then, the vectorized form of \(C({\varvec{X}}, {\varvec{Y}})\) can be defined as

$$\begin{aligned} \begin{aligned} {\varvec{c}}({\varvec{X}}, {\varvec{Y}})&= \mathrm{{vec}} (C({\varvec{X}}, {\varvec{Y}})) \in \mathbb {R}^{nm}\\&= \left[ \varOmega {{\varvec{X}} \atopwithdelims (){\varvec{Y}}} \right] \circ \left[ \varOmega {{\varvec{X}} \atopwithdelims (){\varvec{Y}}} \right] , \end{aligned} \end{aligned}$$
(35)

where \(\varOmega \) is defined as in (5) and the operation \(\circ \) is element-wise product.

The Wasserstein distance By solving LP with the cost vector defined in (35) on the observed data \({\varvec{X}}^\mathrm{{obs}}\) and \({\varvec{Y}}^\mathrm{{obs}}\), we obtain the set of selected basic variables

$$\begin{aligned} \mathcal {M}_\mathrm{{obs}} = \mathcal {M}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}). \end{aligned}$$

Then, the Wasserstein distance can be re-written as (we denote \(W = W(P_n, Q_m)\) for notational simplicity)

$$\begin{aligned} W&= \hat{{\varvec{t}}}^\top {\varvec{c}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}) \\&= \hat{{\varvec{t}}}_{\mathcal {M}_\mathrm{{obs}}}^\top {\varvec{c}}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}) \\&= \hat{{\varvec{t}}}_{\mathcal {M}_\mathrm{{obs}}}^\top \Bigg [ \left[ \varOmega _{\mathcal {M}_\mathrm{{obs}}, :} {{\varvec{X}}^\mathrm{{obs}} \atopwithdelims (){\varvec{Y}}^\mathrm{{obs}}} \right] \circ \left[ \varOmega _{\mathcal {M}_\mathrm{{obs}}, :} {{\varvec{X}}^\mathrm{{obs}} \atopwithdelims (){\varvec{Y}}^\mathrm{{obs}}} \right] \Bigg ]. \end{aligned}$$

Hypothesis testing Our goal is to test the following hypothesis:

$$\begin{aligned} \mathrm{H}_0: \hat{{\varvec{t}}}_{\mathcal {M}_\mathrm{{obs}}}^\top \Bigg [ \left[ \varOmega _{\mathcal {M}_\mathrm{{obs}}, :} { {\varvec{\mu }}_{{\varvec{X}}} \atopwithdelims (){\varvec{\mu }}_{{\varvec{Y}}}} \right] \circ \left[ \varOmega _{\mathcal {M}_\mathrm{{obs}}, :} { {\varvec{\mu }}_{{\varvec{X}}} \atopwithdelims (){\varvec{\mu }}_{{\varvec{Y}}}} \right] \Bigg ] = 0. \end{aligned}$$

Unfortunately, it is technically difficult to directly test the above hypothesis. Therefore, we propose to test the following equivalent one:

$$\begin{aligned}&\mathrm{H}_0: \hat{{\varvec{t}}}_{\mathcal {M}_\mathrm{{obs}}}^\top \left[ \varTheta _{\mathcal {M}_\mathrm{{obs}}, :} { {\varvec{\mu }}_{{\varvec{X}}} \atopwithdelims (){\varvec{\mu }}_{{\varvec{Y}}}} \right] = 0 \\ \Leftrightarrow ~&\mathrm{H}_0: {\varvec{\eta }}^\top { {\varvec{\mu }}_{{\varvec{X}}} \atopwithdelims (){\varvec{\mu }}_{{\varvec{Y}}}} = 0 \end{aligned}$$

where \(\varTheta \) is defined as in (5) and \({\varvec{\eta }}= \varTheta _{\mathcal {M}_\mathrm{{obs}}, :}^\top \hat{{\varvec{t}}}_{\mathcal {M}_\mathrm{{obs}}}\).

Conditional SI To test the aforementioned hypothesis, we consider the following selective p-value:

$$\begin{aligned} p_\mathrm{{selective}} = \mathbb {P}_\mathrm{H_0} \left( \left| {\varvec{\eta }}^\top {{\varvec{X}} \atopwithdelims (){\varvec{Y}}} \right| \ge \left| {\varvec{\eta }}^\top {{\varvec{X}}^\mathrm{{obs}} \atopwithdelims (){\varvec{Y}}^\mathrm{{obs}}} \right| \mid \mathcal {E}\right) \end{aligned}$$

where the conditional selection event is defined as

$$\begin{aligned} \mathcal {E}= \left\{ \begin{array}{l} \mathcal {M}({\varvec{X}}, {\varvec{Y}}) = \mathcal {M}_\mathrm{{obs}}, \\ \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{X}}, {\varvec{Y}}) = \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}) \\ {\varvec{q}}({\varvec{X}}, {\varvec{Y}}) = {\varvec{q}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}) \end{array} \right\} . \end{aligned}$$

Our next task is to identify the conditional data space whose data satisfies \(\mathcal {E}\).

Characterization of the conditional data space Similar to the discussion in Sect. 3, the data are restricted on the line due to the conditioning on the nuisance component \({\varvec{q}}({\varvec{X}}, {\varvec{Y}})\). Then, the conditional data space is defined as

$$\begin{aligned} \mathcal {D}= \Big \{ ({\varvec{X}} ~ {\varvec{Y}})^\top = {\varvec{a}} + {\varvec{b}} z \mid z \in \mathcal {Z}\Big \}, \end{aligned}$$

where

$$\begin{aligned} \mathcal {Z}= \left\{ z \in \mathbb {R}~ \Big | \begin{array}{l} \mathcal {M}({\varvec{a}} + {\varvec{b}} z) = \mathcal {M}_\mathrm{{obs}}, \\ \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{a}} + {\varvec{b}} z) = \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}) \end{array} \right\} . \end{aligned}$$

The remaining task is to construct \(\mathcal {Z}\). We can decompose \(\mathcal {Z}\) into two separate sets as \(\mathcal {Z}= \mathcal {Z}_1 \cap \mathcal {Z}_2\), where

$$\begin{aligned} \mathcal {Z}_1&= \{ z \in \mathbb {R}\mid \mathcal {M}({\varvec{a}} + {\varvec{b}} z) = \mathcal {M}_\mathrm{{obs}}\}, \\ \mathcal {Z}_2&= \{ z \in \mathbb {R}\mid \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{a}} + {\varvec{b}} z) = \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}}) \}. \end{aligned}$$

The construction of \(\mathcal {Z}_2\) is as follows (we denote \({\varvec{s}}_{\mathcal {M}_\mathrm{{obs}}} = \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{X}}^\mathrm{{obs}}, {\varvec{Y}}^\mathrm{{obs}})\) for notational simplicity):

$$\begin{aligned} \mathcal {Z}_2&= \{ z \in \mathbb {R}\mid \mathcal {S}_{\mathcal {M}_\mathrm{{obs}}}({\varvec{a}} + {\varvec{b}} z) = {\varvec{s}}_{\mathcal {M}_\mathrm{{obs}}}\} \\&= \left\{ z \in \mathbb {R}\mid \mathrm{{sign}} \Big (\varOmega _{\mathcal {M}_\mathrm{{obs}}, :} ({\varvec{a}} + {\varvec{b}} z) \Big ) = {\varvec{s}}_{\mathcal {M}_\mathrm{{obs}}} \right\} \\&= \left\{ z \in \mathbb {R}\mid {\varvec{s}}_{\mathcal {M}_\mathrm{{obs}}} \circ \varOmega _{\mathcal {M}_\mathrm{{obs}}, :} ({\varvec{a}} + {\varvec{b}} z) \ge {\varvec{0}}\right\} , \end{aligned}$$

which can be obtained by solving the system of linear inequalities. Next, we present the identification of \(\mathcal {Z}_1\). Because we use squared \(\ell _2\) distance to define the cost matrix, the LP with the parametrized data \({\varvec{a}} + {\varvec{b}} z\) is written as follows:

$$\begin{aligned}&\min \limits _{{\varvec{t}} \in \mathbb {R}^{nm}} ~ {\varvec{t}}^\top [ \varOmega ({\varvec{a}} + {\varvec{b}} z) \circ \varOmega ({\varvec{a}} + {\varvec{b}} z)] ~ ~ \text {s.t.} ~ ~ S {\varvec{t}} = {\varvec{h}}, {\varvec{t}} \ge {\varvec{0}} \\ \Leftrightarrow&\min \limits _{{\varvec{t}} \in \mathbb {R}^{nm}} ~ ({\varvec{u}} + {\varvec{v}} z + {\varvec{w}} z^2)^\top {\varvec{t}} ~ ~ \text {s.t.} ~ ~ S {\varvec{t}} = {\varvec{h}}, {\varvec{t}} \ge {\varvec{0}}, \end{aligned}$$

where

$$\begin{aligned} {\varvec{u}}&= (\varOmega {\varvec{a}}) \circ (\varOmega {\varvec{a}}),\\ {\varvec{v}}&= (\varOmega {\varvec{a}}) \circ (\varOmega {\varvec{b}}) + (\varOmega {\varvec{b}}) \circ (\varOmega {\varvec{a}}), \\ {\varvec{w}}&= (\varOmega {\varvec{b}}) \circ (\varOmega {\varvec{b}}), \end{aligned}$$

and S and \({\varvec{h}}\) are the same as in (7). By fixing \(\mathcal {M}_\mathrm{{obs}}\) as the optimal basic index set, the relative cost vector w.r.t to the set of non-basic variables is defines as

$$\begin{aligned} {\varvec{r}}_{\mathcal {M}^c_\mathrm{{obs}}} = \tilde{{\varvec{u}}} + \tilde{{\varvec{v}}} z + \tilde{{\varvec{w}}} z^2, \end{aligned}$$

where

$$\begin{aligned} \tilde{{\varvec{u}}}&= \left( {\varvec{u}}_{\mathcal {M}^c_\mathrm{{obs}}}^\top - {\varvec{u}}_{\mathcal {M}_\mathrm{{obs}}}^\top S_{:, \mathcal {M}_\mathrm{{obs}}}^{-1} S_{:, \mathcal {M}^c_\mathrm{{obs}}} \right) ^\top , \\ \tilde{{\varvec{v}}}&= \left( {\varvec{v}}_{\mathcal {M}^c_\mathrm{{obs}}}^\top - {\varvec{v}}_{\mathcal {M}_\mathrm{{obs}}}^\top S_{:, \mathcal {M}_\mathrm{{obs}}}^{-1} S_{:, \mathcal {M}^c_\mathrm{{obs}}} \right) ^\top , \\ \text { and } ~~ \tilde{{\varvec{w}}}&= \left( {\varvec{w}}_{\mathcal {M}^c_\mathrm{{obs}}}^\top - {\varvec{w}}_{\mathcal {M}_\mathrm{{obs}}}^\top S_{:, \mathcal {M}_\mathrm{{obs}}}^{-1} S_{:, \mathcal {M}^c_\mathrm{{obs}}} \right) ^\top . \end{aligned}$$

The requirement for \(\mathcal {M}_\mathrm{{obs}}\) to be the optimal basis index set is \({\varvec{r}}_{\mathcal {M}^c_\mathrm{{obs}}} \ge {\varvec{0}}\) (i.e., the cost in minimization problem will never decrease when the non-basic variables become positive and enter the basis). Finally, the set \(\mathcal {Z}_1\) can be defined as

$$\begin{aligned} \mathcal {Z}_1&= \{ z \in \mathbb {R}\mid \mathcal {M}({\varvec{a}} + {\varvec{b}} z) = \mathcal {M}_\mathrm{{obs}}\}, \\ \mathcal {Z}_1&= \{ z \in \mathbb {R}\mid {\varvec{r}}_{\mathcal {M}^c_\mathrm{{obs}}} = \tilde{{\varvec{u}}} + \tilde{{\varvec{v}}} z + \tilde{{\varvec{w}}} z^2 \ge {\varvec{0}}\}, \end{aligned}$$

which can be obtained by solving the system of quadratic inequalities.

B Experiment on high-dimensional data

We generated the dataset \(X = \{{\varvec{x}}_{i, :} \}_{i \in [n]}\) with \({\varvec{x}}_{i, :} \sim \mathbb {N}({\varvec{1}}_d, I_d)\) and \(Y = \{{\varvec{y}}_{j, :} \}_{j \in [m]}\) with \({\varvec{y}}_{j, :} \sim \mathbb {N}({\varvec{1}}_d + \varDelta , I_d)\) (element-wise addition). We set \(n = m = 20, \varDelta = 2\), and ran 10 trials for each \(d \in \{100, 150, 200, 250\}\). The results in Fig. 10 show that the proposed method still has reasonable computational time.

Fig. 10
figure 10

Computational time of the proposed method when increasing the dimension d

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Duy, V.N.L., Takeuchi, I. Exact statistical inference for the Wasserstein distance by selective inference. Ann Inst Stat Math 75, 127–157 (2023). https://doi.org/10.1007/s10463-022-00837-3

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