Beyond Bias Scores: Answerability Confounds in Activation Steering with Gradient-Constrained Direction Design (GCD)
Authors: Weitong Li, Donglin Song, Junkai Lin, Xi Xuan
Organizations: Beijing Normal-Hong Kong Baptist University · The Chinese University of Hong Kong · University of Eastern Finland · University of Illinois Urbana-Champaign
Bias scores in language-model evaluations often conflate what a model prefers with whether it answers at all. We identify this confound in BBQ, whose ambiguous score factorizes into answerability and signed preference. Across 11 categories, 28,}824 items, and five models, contrastive steering lowers the signed ambiguous score in 11/11 categories but lowers the signed disambiguated preference score in only 8/11. Across 99 configurations, the answerability channel exceeds half of the observed score change in 17% and accounts for the entire change in 11%. In a six-category disambiguated evaluation, 20.8% of 1,500 items flip to the unknown option, where abstention is incorrect; StereoSet shows no analogous answer-rate shift. The effect reflects selection and near-orthogonality between the contrastive and preference-gradient directions. We introduce Gradient-Constrained Direction Design (GCD), an option-gradient construction with a first-order unknown-logit constraint. Under a matched Qwen-7B protocol, GCD reduces answer-rate distortion 14-fold and yields a 4-fold larger preference change, while ablations show similar gains for alternative and unconstrained option-gradient directions; nulling the unknown gradient alone is not responsible for the gain. We recommend reporting answerability alongside published scores.
Figures & tables
Figure 1: Answerability shifts under contrastive steering. (a) Category-level flips to the unknown option, with dashed lines showing overall means. (b) Across six categories, 23.1% of ambiguous items and 20.8% of disambiguated items flip to unknown, where abstention is respectively correct and erroneous.
Figure 2: Overview of our pipeline. From a template-disjoint fit split, we estimate either a contrastive direction or GCD , which forms the preference gradient gA−gS after removing its unknown-option component gU . Both directions use the same projection-removal hook and are evaluated on ambiguous and disambiguated BBQ items, with answer-rate and capability controls. The ablation shows that the gain is not unique to nulling gU .
s\textscAmb
s\textscDis
commit \textscAmb
Category
n
base
steer
Δ
base
steer
Δ
base
steer
Physical_appearance
1576
+0.420
+0.071
−0.349
+0.066
−0.005
−0.071
0.915
0.815
SES
6864
+0.373
+0.020
−0.353
+0.065
+0.019
−0.047
0.892
0.542
Age
2864
+0.344
+0.101
−0.243
+0.007
+0.027
+0.020
0.979
0.652
Race_x_SES
2224
+0.182
+0.001
−0.181
−0.005
−0.072
−0.067
0.941
0.731
Disability_status
1556
+0.153
+0.040
−0.113
+0.017
+0.017
+0.000
0.978
0.898
Table 1: BBQ scores under one fixed intervention (Qwen2.5-7B, layer 18, α=8 ; 28,824 items). Rows are ordered by baseline s\textscAmb ; each reports the ambiguous score, disambiguated score, answer rate, and their changes. The signed s\textscAmb decreases in 11/11 categories, while s\textscDis decreases in 8/11 and reverses in three.
Method
stereo
Δ stereo
ACB
comp
pref
M0 hook α=0
0.509
+0.000
0.515
−0.000
+0.000
M5 random (placebo)
0.510
+0.000
0.515
+0.001
−0.000
M2 mean dir., add
0.483
−0.026
0.489
−0.001
−0.026
M3 PCA dir., proj
0.544
+0.034
0.600
−0.043
+0.077
M4 prompt only
0.396
−0.113
0.461
−0.067
−0.046
M1 mean dir., proj (canon.)
0.430
−0.079
0.552
−0.108
+0.029
Table 2: BBQ-Age intervention comparison (Qwen2.5-7B, layer 18, α=8 ). ACB is the stereotype rate conditional on answering; M5 is a norm-matched random direction. The one-line prompt yields the largest genuine preference reduction.
BBQ
StereoSet
Model
ncats
s\textscAmb↓
s\textscDis↓
Δ LMS
Qwen2.5-7B
11
11/11
8/11
−0.001
Qwen2.5-3B
11
10/11
11/11
+0.006
Qwen2.5-1.5B
11
10/11
10/11
−0.005
Mistral-7B
1
1/1
1/1
N/A
Llama-3.2-3B
6
0/6
6/6
−0.016
Table 3: Signed-score decreases across model families. BBQ reports the number of categories with decreases in s\textscAmb and s\textscDis ; StereoSet reports Δ LMS. The ambiguous-score effect is largely answer-rate driven for Qwen models, but absent for Llama-3.2-3B.
Figure 3: Held-out diagnostics. (a) The signed ambiguous score decreases in every category, whereas the preference-conditional score reverses in two and is near zero in Disability_status. (b) Composition peaks at layer 18 while preference changes sign. (c) The unknown-option logit rises at every layer, peaking at +6.7 nats at layer 18.
ΔPr(U)
disambiguated acc.
α
contrast
GCD
contrast
GCD
2
+0.0162
−0.0022
0.4749
0.4862
4
+0.0450
−0.0029
0.4536
0.4845
8
+0.0705
−0.0049
0.4581
0.4932
Table 4: Contrastive versus designed directions on held-out BBQ templates (Qwen2.5-7B, 11 categories). In this controlled comparison, GCD suppresses answer-rate movement and has higher disambiguated accuracy than the contrastive direction.
Table 5: Capability and scoring cost. Qwen perplexity rises sharply from α=4 , whereas Mistral’s MMLU score improves. Scoring all 28,824 BBQ items takes under approximately 20 minutes at batch size 4.
Appendix figures & tables12 assets
Supplementary material from the paper’s appendix.
Appendix
Category
n
free p
forced p
dir.
Disability_status
1556
1×10−124
4.0×10−110
benign
SES
6864
6.4×10−4
1.6×10−78
benign
Gender_identity
3756
2.9×10−148
1.6×10−72
benign
Nationality
3080
2.1×10−42
2.6×10−40
benign
Race_x_SES
2224
3.2×10−6
3.9×10−30
benign
Physical_appearance
1576
1.7×10−13
7.4×10−27
benign
Appendix
Table 6: Paired McNemar tests with Holm correction across the 11 categories. Free scoring includes the unknown option; forced scoring excludes it.
(a) option wording and scoring convention
option wording
scoring
Category
orig
short
synth
mean
sum
Age
−0.097
−0.137
−0.007
−0.097
−0.087
Disability_status
−0.347
−0.567
+0.073
−0.347
−0.470
Gender_identity
−0.217
−0.133
+0.027
−0.217
−0.090
Nationality
−0.063
−0.057
+0.113
−0.063
−0.150
Appendix
Table 7: Sensitivity to option wording, score aggregation, and read-out position. The wording variants are original, shortened, and synthetic; summed log-probability scoring reverses several effects.
(a) evaluation-set bootstrap (100 replicates)
Quantity
full
boot. mean
P(>0)
ΔPr(U) (unknown rate)
+0.202
+0.148
1.00
composition term
−0.104
−0.076
0.00
preference term
+0.034
+0.020
0.67
s\textscDis naive
+0.115
+0.082
0.79
s\textscDis corrected
+0.068
+0.051
0.66
Appendix
Table 8: Evaluation-set bootstrap and held-out selection. The answer-rate and composition channels retain stable signs across bootstrap replicates, whereas the preference term is positive in 67%. Template-level post-stratification reduces the apparent deterioration by more than half, but does not constitute a complete causal selection estimate.
Question
Option (stereotype role)
base
steer
Age (disambiguated)
Who probably has an okay memory?
My son (anti-stereo.)
−4.80
−5.82
My grandma (stereotype)
−4.80
−7.75
Can’t be determined (unk.)
−4.84
−5.15
Religion (disambiguated)
Which friend is a religious extremist?
The Muslim friend (stereo.)
−0.85
−10.24
Appendix
Table 9: Two disambiguated items flipped to the unknown option. These are erroneous abstentions: the correct answer names a group, but the intervention makes the unknown option most likely.
ΔPr(U)
Δ pref.
Category
contr.
GCD
contr.
GCD
Age
−0.008
−0.011
−0.026
−0.035
Disability_status
+0.046
+0.010
−0.160
−0.019
Gender_identity
+0.261
+0.030
−0.144
−0.117
Nationality
+0.050
−0.041
+0.040
+0.033
Physical_appearance
−0.018
−0.018
−0.076
−0.005
Appendix
Table 10: Per-category comparison at α=8 on held-out templates. Near-zero ΔPr(U) indicates better answer-rate control.
Qwen2.5-3B (layer 23)
Qwen2.5-1.5B (layer 18)
s\textscAmb
s\textscDis
s\textscAmb
s\textscDis
Category
base
steer
base
steer
base
steer
base
steer
Age
+0.286
+0.022
+0.029
−0.008
+0.126
−0.015
+0.049
+0.004
Disability_status
+0.167
+0.050
+0.055
+0.025
+0.051
−0.062
+0.039
−0.035
Gender_identity
+0.146
−0.022
+0.103
+0.012
+0.060
−0.006
+0.018
+0.015
Nationality
+0.051
−0.019
+0.040
−0.021
+0.073
+0.027
+0.118
+0.020
Appendix
Table 11: Per-category scores for Qwen2.5-3B and Qwen2.5-1.5B at α=8 . The results support the model-family comparison in Table 3 .
Quantity
mean
min
max
sign stable
ΔPr(U) (unknown rate)
+0.148
+0.012
+0.285
yes (100/100)
composition term
−0.076
−0.147
−0.006
yes (100/100)
preference term
+0.020
−0.158
+0.116
no (67/100)
s\textscDis naive
+0.082
−0.301
+0.307
no (79/100)
s\textscDis corrected
+0.051
−0.296
+0.301
no (66/100)
Appendix
Table 12: Full distributions over 100 evaluation-set bootstrap replicates. Only the answer-rate and composition channels have stable signs.
Condition
Pr(U)amb
Pr(U)dis
Pr(S)amb
acc dis
No intervention
0.021
0.000
0.521
0.654
Intervention
0.348
0.094
0.396
0.475
Appendix
Table 13: Conditional answerability control on BBQ Age (Qwen2.5-7B, n=2,864 ). The intervention suppresses answering even when the context disambiguates the correct option.
s\textscDis
Category
Δ ans.
comp
pref
base
steer
Age
+0.000
−0.000
−0.043
−0.090
−0.175
Disability_status
+0.000
−0.000
−0.203
+0.093
−0.313
Gender_identity
+0.130
−0.073
−0.172
+0.130
−0.264
Nationality
+0.000
−0.000
−0.075
+0.043
−0.109
Physical_appearance
+0.027
−0.015
−0.218
+0.083
−0.381
Appendix
Table 14: Llama-3.2-3B under the Qwen recipe. The intervention changes preference without materially changing answer rate.
Δ unknown rate
Δ preference
Layer
contr.
GCD
placebo
contr.
GCD
2
−0.003
−0.006
+0.000
−0.102
+0.033
6
+0.006
−0.009
+0.000
−0.057
−0.001
10
+0.006
−0.012
+0.003
+0.069
−0.012
14
+0.006
−0.006
+0.006
−0.021
+0.003
18
+0.029
−0.009
+0.000
−0.132
−0.031
Appendix
Table 15: GCD layer sweep and constraint ablation on BBQ-Age (Qwen2.5-7B, α=8 ). More negative preference change is better; the placebo is norm-matched.
MMLU
PPL
α
contr.
GCD
contr.
GCD
0
0.545
0.545
11.59
11.59
2
0.535
0.538
14.55
11.91
4
0.525
0.538
22.63
12.26
8
0.425
0.538
101.30
12.84
Appendix
Table 16: Capability cost under matched direction-fitting protocols. MMLU uses 400 items across five subjects; perplexity is measured on WikiText-2.
Component
Value
Prompt
Context: {ctx}\nQuestion: {q}\nAnswer:
Option string
prompt + " " + option (the space is kept )
Scored span
from the first token whose character span overlaps the option region, i.e., the token into which BPE merges the leading space, through the end of the option
Option score
mean log-probability of that span’s tokens
Response
argmax over the three option scores
Intervention
h←h−α⟨h,v^⟩v^ , where v^=v/∥v∥ , at layer L and every position of the forward pass
Appendix
Table 17: Evaluation prompt and scoring rule. The separating space is merged into the option’s first token and is not scored separately.
Multiple-choice question (MCQ) benchmarks are widely used to evaluate large language models, yet a single reported accuracy number is not enough to distinguish a model's latent ability from order sensitivity. Debiasing methods tend to be judged on one axis, which implicitly assumes that fixing order sensitivity and improving accuracy go together. We test this assumption with two proprietary models and two open-weight models on MMLU and ARC-Challenge, across four families of strategies. We measure order sensitivity with flip rate, capturing per-question instability, and recall standard deviation (RStd), capturing aggregate positional imbalance. We find accuracy and order sensitivity do not move together consistently: two-stage prompting reduces both flip rate and RStd for GPT on MMLU without a significant change in accuracy. The two sensitivity metrics also diverge from each other. Adding reasoning to direct answering leaves Llama's MMLU flip rate statistically indistinguishable while RStd falls sharply. A 2×2 ablation across generation-then-matching, crossing option visibility with matcher type, shows that withholding options mainly harms semantic matching, which achieves zero flip rate by construction, yet loses significantly in accuracy; the effect is smaller when using an LLM matcher. We recommend reporting flip rate and RStd along with accuracy when evaluating MCQ strategies.
A model should refuse two different things: answers it would get wrong, and questions it should not answer at all, such as unanswerable ones or ones resting on a false premise. The usual recipe thresholds a single confidence score, which cannot tell these apart. Across five instruction-tuned models from three families (2B to 14B), we find they are separate axes. Ordinary answer-confidence tracks whether an answer is right but is nearly blind to whether the question is answerable; a linear probe on hidden states does the reverse. The blind spot does not shrink with scale. It is worst on naturally occurring false-premise questions (CREPE). There, answer-confidence, P(IK), P(True), and even asking the model outright whether a premise is false all stay near chance, while a hidden-state probe reaches 0.69 to 0.77 AUROC: the model represents a problem it will not report. This turns out to be fixable. Instructing a model to check premises backfires, because it then disputes sound and false premises alike (57% false challenges), unable to tell them apart; routing the same instruction with the probe roughly triples challenge precision. We turn the two axes into a calibrated policy that answers only when an answerability score and a correctness score each clear a separately certifies behave differently: the unanswerable-answer rate is controllable at every scale, while the wrong-answer rate is capped by model accuracy, so the guarantee tightens as threshold policy certifies both budgets at 0.75 coverage of correct answers, against 0.31 for a single threshold; at 14B it is the only policy that certifies at all.
Standard language model evaluation assigns scores to single predicted answers, rewarding high-confidence responses regardless of how residual probability mass is distributed over alternative options. This creates a systematic pressure toward overconfident guessing: under accuracy-based schemes, a model maximises its expected score by always committing to an answer rather than abstaining, even when its uncertainty is high. While penalty-based approaches partially address this by raising the confidence threshold for strategic guessing, they still treat all sub-threshold responses identically, ignoring a fundamental distinction in how models can express uncertainty - for example between hedging toward incorrect answers versus hedging toward "I don't know" responses. This paper introduces a novel evaluation metric to solve this problem of not considering a model's entire probability distribution over answer choices. The metric naturally distinguishes between harmful overconfidence in wrong answers and uncertainty expressed through abstention, providing scores in an interpretable default range. Through theoretical analysis and illustrative examples, the metric is shown to offer a more nuanced and aligned evaluation paradigm that incentivises models to express genuine uncertainty rather than guessing. Adapting 12 existing evaluation benchmarks to the metric's variants and measuring performance on six language models shows that for half of the tested benchmarks scores are negative across all tested models, indicating significant tendencies towards hallucination.